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\begin{document}

\begin{frontmatter}
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\title{Generalized Multilinear Models for Sufficient Dimension Reduction on Tensor-valued Predictors}
\runtitle{GMLM for SDR on tensor-valued Predictors}

\begin{aug}
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\author[A]{\fnms{Daniel}~\snm{Kapla}\ead[label=e1]{daniel.kapla@tuwien.ac.at}}
\and
\author[A]{\fnms{Efstathia}~\snm{Bura}\ead[label=e2]{efstathia.bura@tuwien.ac.at}}
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\address[A]{Institute of Statistics and Mathematical Methods in Economics, Faculty of Mathematics and Geoinformation, TU Wien \\
\printead[presep={\ }]{e1,e2}}
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\begin{abstract}
We consider supervised learning problems with tensor-valued input. We derive multi-linear sufficient reductions for the regression or classification problem by modeling the conditional distribution of the predictors given the response as a member of the quadratic exponential family. We develop estimation procedures of sufficient reductions for both continuous and binary tensor-valued predictors. We prove the consistency and asymptotic normality of the estimated sufficient reduction using manifold theory. For multi-linear normal predictors, the estimation algorithm is highly computationally efficient and is also applicable to situations where the dimension of the reduction exceeds the sample size. Our method outperforms competing techniques in both simulated settings and real-world datasets involving continuous and binary tensor-valued predictors. 
\end{abstract}

%%% see: https://mathscinet.ams.org/mathscinet/msc/msc2020.html
%% 62B05 - Sufficient statistics and fields
%% 62E20 - Asymptotic distribution theory in statistics
%% 62F12 - Asymptotic properties of parametric estimators
%% 62F30 - Parametric inference under constraints
%% 62H12 - Estimation in multivariate analysis

\begin{keyword}[class=MSC]
\kwd[Primary ]{62E20}
\kwd{62J05, 62F12}
\kwd[; secondary ]{62F30}
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\begin{keyword}
\kwd{Regression}
\kwd{Asymptotics}
\kwd{Manifold Theory}
\kwd{Constrained Optimization}
\kwd{Maximum Likelihood Estimation}
\end{keyword}

\end{frontmatter}

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\section{Introduction}\label{sec:introduction}
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Complex data are collected at different times and/or under conditions often involving a large number of multi-indexed variables represented as tensor-valued data \cite[]{KoldaBader2009}. A tensor is a multi-dimensional array of numbers. As such, \textit{Tensors}  are a mathematical tool to represent data of complex structure in that they are a generalization of matrices to higher dimensions. They occur in large-scale longitudinal studies \cite[e.g.][]{Hoff2015}, in agricultural experiments, chemometrics and spectroscopy \cite[e.g.][]{LeurgansRoss1992,Burdick1995}, in signal and video processing where sensors produce multi-indexed data, e.g. over spatial, frequency, and temporal dimensions \cite[e.g.][]{DeLathauwerCastaing2007,KofidisRegalia2001}, and in telecommunications \cite[e.g.][]{DeAlmeidaEtAl2007}. Other examples of multiway data include 3D images of the brain, where the modes are the 3 spatial dimensions, and spatio-temporal weather imaging data, a set of image sequences represented as 2 spatial modes and 1 temporal mode.

Tensor regression models have been proposed to leverage the structure inherent in tensor-valued data. For instance, \cite{HaoEtAl2021,ZhouLiZhu2013} focus on tensor covariates, while \cite{RabusseauKadri2016,LiZhang2017,ZhouLiZhu2013} focus on tensor responses, and \cite{Hoff2015,Lock2018} consider tensor on tensor regression. \cite{HaoEtAl2021} modeled a scalar response as a flexible nonparametric function of tensor covariates. \cite{ZhouLiZhu2013} assume the scalar response has a distribution in the exponential family given the tensor-valued predictors with the link modeled as a multi-linear function of the predictors. \cite{RabusseauKadri2016} model the tensor-valued response as a linear model with tensor-valued regression coefficients subject to a multi-linear rank constraint. \cite{LiZhang2017} approach the problem with a similar linear model but instead of a low-rank constraint the error term is assumed to have a separable Kronecker product structure while using a generalization of the envelope model \cite[]{CookLiChiaromonte2010}. \cite{ZhouEtAl2023} focus on partially observed tensor response given vector-valued predictors with mode-wise sparsity constraints in the regression coefficients. \cite{Hoff2015} extends an existing bilinear regression model to a tensor on tensor of conformable modes and dimensions regression model based on a Tucker product. \cite{Lock2018} uses a tensor contraction to build a penalized least squares model for a tensor with an arbitrary number of modes and dimensions.

We consider the general regression problem of fitting a response of general form (univariate, multivariate, tensor-valued) on a tensor-valued predictor. We operate in the context of \textit{sufficient dimension reduction} (SDR) \cite[e.g.][]{Cook1998,Li2018} based on inverse regression, which leads to regressing the tensor-valued predictor on tensor-valued functions of the response (tensor on tensor regression). %In our setting, this necessitates transforming the response to tensor-valued functions, regardless of whether it is itself tensor-valued. 
Ours is a \textit{model-based} SDR method for tensor-valued data, where the conditional distribution of the inverse predictors ($\ten{X}$) belongs to a parametric family whose parameter depends on the response of the forward regression of $Y$ on $\ten{X}$. Specifically, we assume the conditional distribution of the tensor-valued predictors given the response belongs to the quadratic exponential family for which the first and second moments admit a separable Kronecker product structure. The quadratic exponential family contains the multi-linear normal and the multi-linear Ising distributions for continuous and binary tensor-valued random variables, respectively. 

%In this paper, we present a model-based \emph{Sufficient Dimension Reduction} (SDR) method for tensor-valued data with distribution in  
%Because of the setting, our method shares commonalities with the tensor regression models referred to above, yet the modeling and methodology are novel. 
   %We bypass the explosion of the number of parameters to estimate by assuming the inverse regression error covariance has Kronecker product structure.% as do \cite{LiZhang2017}. 
We obtain the maximum dimension reduction of the tensor-valued predictor without losing any information about the response. We derive maximum likelihood  estimates of the sufficient reduction that enjoy the attendant optimality properties, such as consistency, efficiency and asymptotic normality. 
The main challenge in estimation in multi-linear tensor regression models is the non-identifiability of the parameters, as has been acknowledged by researchers in this field (e.g., \cite{LiZhang2017, Lock2018}). In contrast to past approaches (e.g., \cite{ZhouLiZhu2013,Hoff2015, SunLi2017, LiZhang2017, LlosaMaitra2023}), ours does not require any penalty terms and/or sparsity constraints in order to address this issue.  %In the case of a tensor (multi-linear) normal, given the tensor-valued function of the response, our model exhibits similarities to the multi-linear modeling of \cite{Hoff2015}, but we use a generalized multi-linear model and estimate the parameters with maximum likelihood instead of least squares. 
Instead, we model the parameter space as a smooth embedded manifold to ensure identifiability while enabling rich modeling flexibility.

Our main contributions are (i) formulating the dimension reduction problem via the quadratic exponential family modeling up to two-way interaction that allows us to derive the minimal sufficient dimension reduction in closed form, (ii) great flexibility in modeling by defining the parameter space as a smooth embedded manifold, (iii) deriving the maximum likelihood estimator of the sufficient reduction subject to multi-linear constraints and overcoming parameter non-identifiability, (iv) establishing the consistency and asymptotic normality of the estimators, and (v) developing computationally efficient estimation algorithms (very fast in the case of multi-linear normal predictors).

We clarify our approach and highlight our contributions via a simple example where the predictor is matrix-valued (tensor of order $2$) and the response is univariate (binary). The electroencephalography (EEG) data set\footnote{\textsc{Begleiter, H.} (1999). EEG Database. \textit{Neurodynamics Laboratory, State University of New York Health Center} Donated by Lester Ingber. \url{http://kdd.ics.uci.edu/databases/eeg/eeg.data.html}} of $77$ alcoholic and $45$ control subjects is commonly used in EEG signal analysis. EEG is a noninvasive neuroimaging technique that involves the placement of electrodes on the scalp to record electrical activity of the brain. Each subject data point consists of a $p_1\times p_2 = 256\times 64$ matrix, with each row representing a time point and each column a channel. The measurements were obtained by exposing each individual to visual stimuli and measuring voltage values from $64$ electrodes placed on the subjects' scalps sampled at $256$ time points over $1$ second ($256$ Hz). For each subject $120$ trials were measured under different stimulus conditions. To contrast our approach to previous methods applied to this data set (e.g., \cite{LiKimAltman2010,PfeifferForzaniBura2012,DingCook2015,PfeifferKaplaBura2021}), we use a single stimulus condition (S1), and for each subject, we average over all the trials under this condition. The data are $(\ten{X}_i, Y_i)$, $i = 1, \ldots, 122$, where $\ten{X}_i$ is a $256\times 64$ matrix (two-mode tensor), with each entry representing the mean voltage value of subject $i$ at a combination of a time point and a channel, averaged over all trials under the S1 stimulus condition, and $Y$ is a binary outcome variable with $Y_i = 1$ for an alcoholic and $Y_i = 0$ for a control subject.

We assume $\ten{X} \mid Y$ follows the bilinear model
\begin{equation}\label{eq:bilinear}
    \ten{X} = \bar{\mat{\eta}} + \mat{\alpha}_1\mat{F_Y}\t{\mat{\alpha}_2} + \mat{\epsilon}
\end{equation}
where $\mat{F}_Y\equiv Y$ is a $1\times 1$ matrix (in general, $\mat{F}_Y$ is a tensor-valued function of $Y$), $\vec{\mat{\epsilon}}\sim\mathcal{N}(\mat{0}, \mat{\Sigma}_2\otimes\mat{\Sigma}_1)$, and $\otimes$ signifies the Kronecker product operator. Model \eqref{eq:bilinear} expresses that $\E(\ten{X}\mid Y) = \bar{\mat{\eta}} + \mat{\alpha}_1\mat{F}_Y\t{\mat{\alpha}_2}$ solely contains the information in $Y$ about $\ten{X}$. Alternatively, \eqref{eq:bilinear} can be written as $\ten{X} = \bar{\mat{\eta}} + \mat{\alpha}_1 \mat{F_Y}\t{\mat{\alpha}_2} + \mat{\Sigma}_1^{-1/2}\mat{U}\mat{\Sigma}_2^{-1/2}$, where $\vec{\mat{U}}\sim\mathcal{N}(\mat{0}, \mat{I}_{p_2}\otimes\mat{I}_{p_1})$. Thus, $\ten{X}$ follows a matrix normal distribution with mean $\bar{\mat{\eta}} + \mat{\alpha}_1 \mat{F_Y}\t{\mat{\alpha}_2}$ and variance-covariance structure $\mat{\Sigma}_2\otimes\mat{\Sigma}_1$. We derive the \textit{minimal sufficient reduction} of the predictor $\ten{X}$ for the regression (classification) of $Y$ on $\ten{X}$ to be
\begin{displaymath}
    \ten{R}(\ten{X})
    = \t{(\mat{\Sigma}_1^{-1}\mat{\alpha}_1)}(\ten{X}-\bar{\mat{\eta}})(\mat{\Sigma}_2^{-1}\mat{\alpha}_2)
    = \t{\mat{\beta}_1}(\ten{X}-\bar{\mat{\eta}})\mat{\beta}_2.
\end{displaymath}
Our approach involves optimization of the likelihood of $\ten{X}$ conditional on $Y$ subject to the non-convex nonlinear constraint that both the mean and covariance parameter spaces have Kronecker product structure, where the components of the Kronecker product are non-identifiable. This is resolved by letting the parameter space be a smooth embedded submanifold. We derive the maximum likelihood estimator of the minimal sufficient reduction and use it to predict the binary response. Our method exhibits uniformly better classification accuracy when compared to competing existing techniques (see Table \ref{tab:eeg}). A key advantage of our approach is that it circumvents the dimensionality challenge ($p_1 \times p_2 = 256 \times 64 = 16384 \gg 122 = n$) without having to artificially reduce the dimension by preprocessing the data, as in \cite{LiKimAltman2010, PfeifferForzaniBura2012, PfeifferKaplaBura2021, DingCook2015}, or imposing simplifying assumptions and/or sparsity or regularization constraints \cite[]{PfeifferKaplaBura2021,LiZhang2017,SunLi2017, Lock2018}.

Even though our motivation is rooted in the SDR perspective, our inverse regression approach also  applies to any regression model with a tensor-valued response and predictors of any type. Thus, it can be used as a stand-alone model for such data regardless of whether one is interested in deriving sufficient reductions or simply regressing tensor-valued variables on any type of predictor. Our tensor-to-tensor inverse regression model is a generalized multi-linear model that shares a similar structural form with the generalized linear model of \cite{ZhouLiZhu2013, Hoff2015} and \cite{SunLi2017}, but beyond that, the resemblance does not extend further. In effect, our model can be viewed as an extension of reduced rank regression for tensor-valued response regression.
%Our consistency and asymptotic normality results in the framework of the quadratic exponential family for tensor-valued variables apply to both multi-linear normal \cite[]{KolloVonRosen2005,Hoff2011,OhlsonEtAl2013} and multi-linear Ising models, as defined in \cref{sec:ising_estimation}.

The structure of this paper is as follows. We begin by introducing notation in \cref{sec:notation}, followed by a formal definition of the problem in \Cref{sec:problem-formulation}. The proposed model is specified in \cref{sec:gmlm-model}.
\Cref{sec:manifolds} provides a brief introduction to manifolds, serving as the basis for the consistency and asymptotic normality results detailed in \cref{sec:statprop}. A general maximum likelihood estimation procedure is presented and we derive specialized methods for the multi-linear normal and multi-linear Ising distributions in \cref{sec:ml-estimation}.
Simulations for continuous and binary tensor-valued predictors are carried out in \cref{sec:simulations}. We apply our model to EEG data, where the predictor takes the form of two- and three-dimensional arrays, as presented in \cref{sec:data-analysis}.
Finally, we summarize our contributions and highlight potential directions for future research in \cref{sec:discussion}.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Notation}\label{sec:notation}
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Capital Latin letters in \LaTeX{} fraktur font will denote tensors, and Latin and Greek letters in boldface will denote vectors and matrices throughout the paper. A multiway array $\ten{A}$ of dimension $q_1\times \ldots\times q_r$ is a tensor of order\footnote{Also referred to as rank, hence the variable name $r$. We refrain from using this term to avoid confusion with the rank of a matrix.} $r$, in short $r$-tensor, where $r\in\mathbb{N}$ is the number of its modes or axes (dimensions), and we denote it by $\ten{A}\in\mathbb{R}^{q_1\times \ldots\times q_r}$ where $\ten{A}_{i_1,...,i_r} \in \mathbb{R}$ is its $(i_1, \ldots, i_r)$th entry. For example, a $p \times q$ matrix $\mat{B}$ is a tensor of \textit{order} 2 as it has two modes, the rows and columns.

The $k$-mode product $\ten{A}\times_k\mat{B}$ of a tensor $\ten{A}\in\mathbb{R}^{q_1\times q_2\times\cdots\times q_r}$ with a matrix $\mat{B}\in\mathbb{R}^{p\times q_k}$ is a generalization of matrix multiplication that operates along the $k$-th mode of the tensor, summing over the corresponding index $i_k$ to produce a new tensor with appropriately adjusted dimensions:
\begin{displaymath}
    (\ten{A}\times_k\mat{B})_{i_1, \ldots, i_{k-1}, j, i_{k+1}, \ldots, i_r}
    = \sum_{i_k = 1}^{q_k} \ten{A}_{i_1, \ldots, i_r}\ten{B}_{j, i_k}.
\end{displaymath}
The notation $\ten{A}\mlm_{k\in S}\mat{B}_k$ is shorthand for the iterative application of the mode product with matching matrices $\mat{B}_k$, for all indices $k\in S\subseteq\{1, \ldots, r\}$. For example $\ten{A}\mlm_{k\in\{2, 5\}}\mat{B}_k = \ten{A}\times_2\mat{B}_2\times_5\mat{B}_5$. By only allowing $S$ to be a set, this notation is unambiguous because the mode product commutes for different modes; i.e., $\ten{A}\times_j\mat{B}_j\times_k\mat{B}_k = \ten{A}\times_k\mat{B}_k\times_j\mat{B}_j$ for $j\neq k$. For matrices $\mat{A}, \mat{B}_1, \mat{B}_2$, the relation to classic matrix-matrix multiplication is
\begin{displaymath}
    \mat{A}\times_1\mat{B}_1 = \mat{B}_1\mat{A}, \qquad
    \mat{A}\times_2\mat{B}_2 = \mat{A}\t{\mat{B}_2}, \qquad
    \mat{A}\mlm_{k = 1}^2\mat{B}_k = \mat{A}\mlm_{k \in \{1, 2\}}\mat{B}_k = \mat{B}_1\mat{A}\t{\mat{B}_2}.
\end{displaymath}
An application of the mode product over all indices is also known as the \emph{multi-linear multiplication}, or \emph{Tucker operator} \cite[]{Kolda2006,KofidisRegalia2001}.

The \emph{vectorization} $\vec(\ten{A})$ of a tensor $\ten{A}$ is a vector consisting of the same elements as $\ten{A}$ arranged in lexicographic index order. We use the notation $\ten{A}\equiv \ten{B}$ if and only if $\vec(\ten{A}) = \vec(\ten{B})$. The \emph{flattening} (or \emph{unfolding} or \emph{matricization}) of $\ten{A}$ along mode $k$, denoted as $\ten{A}_{(k)}$, is defined as the matrix obtained by rearranging the elements of $\ten{A}$ such that the $k$-th mode is represented as the columns of the resulting matrix (see \cref{app:examples} for examples.)

The \emph{inner product} between two tensors with the same number of elements is $\langle\ten{A}, \ten{B}\rangle = \t{(\vec\ten{A})}(\vec\ten{B})\in\mathbb{R}$ (allowing tensors of different order), and the \emph{Frobenius norm} for tensors is $\|\ten{A}\|_F = \sqrt{\langle\ten{A}, \ten{A}\rangle}$. The \emph{outer product} between two tensors $\ten{A}$ of dimensions $q_1, \ldots, q_r$ and $\ten{B}$ of dimensions $p_1, \ldots, p_l$ is a tensor $\ten{A}\circ\ten{B}$ of order $r + l$ and dimensions $q_1, \ldots, q_r, p_1, \ldots, p_l$, such that $\ten{A}\circ\ten{B} \equiv (\vec\ten{A})\t{(\vec{\ten{B}})}$. The iterated outer and iterated Kronecker products are written as
\begin{displaymath}
    \bigouter_{k = 1}^{r} \mat{A}_k = \mat{A}_1\circ\ldots \circ\mat{A}_r,
    \qquad
    \bigkron_{k = 1}^{r} \mat{A}_k = \mat{A}_1\otimes\ldots \otimes\mat{A}_r
\end{displaymath}
where the order of iteration is important. Similar to the outer product, we extend the Kronecker product to $r$-tensors $\ten{A}, \ten{B}$ yielding a $2r$-tensor $\ten{A}\otimes\ten{B}$.

Finally, the gradient of a function $\ten{F}(\ten{X})$ of any shape, univariate, multivariate, or tensor-valued, with argument $\ten{X}$ of any shape is defined to be the $p\times q$ matrix 
\begin{displaymath}
    \nabla_{\ten{X}}\ten{F} = \frac{\partial\t{(\vec\ten{F}(\ten{X}))}}{\partial(\vec\ten{X})},
\end{displaymath}
where %the vectorized quantities 
$\vec{\ten{X}}\in\mathbb{R}^p$ and $\vec\ten{F}(\ten{X})\in\mathbb{R}^q$. This is consistent with the gradient of a real-valued function $f(\mat{x})$ where $\mat{x}\in\mathbb{R}^p$ as $\nabla_{\mat{x}}f\in\mathbb{R}^{p\times 1}$ \cite[][Ch.~15]{Harville1997}.



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\section{Problem Formulation}\label{sec:problem-formulation}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
Our goal is to reduce the complexity of inferring the cumulative distribution function (cdf) $F$ of $Y\mid \ten{X}$, where $\ten{X}$ is assumed to admit $r$-tensor structure of dimension $p_1\times ... \times p_r$ with continuous or discrete entries, and the response  $Y$ is unconstrained. To this end, %The predictor $\ten{X}$ is a complex object; to simplify the problem 
we assume there exists a tensor-valued function of lower dimension $\ten{R}:\ten{X}\mapsto \ten{R}(\ten{X})$ such that 
\begin{displaymath}
    F(Y\mid \ten{X}) = F(Y\mid \ten{R}(\ten{X})).
\end{displaymath}
where the tensor-valued $\ten{R}(\ten{X})$ has dimension  $q_1\times...\times q_r$ with $q_j\leq p_j$, $j = 1, ..., r$, which represents a dimension reduction along every mode of $\ten{X}$.  Since $\ten{R}(\ten{X})$ replaces the predictors without changing the conditional cdf of $Y\mid \ten{X}$, it is a \emph{sufficient reduction} for the regression $Y\mid\ten{X}$. This formulation is flexible as it allows, for example, to select ``important'' modes by reducing ``unimportant'' modes to be $1$ dimensional.

To find such a reduction $\ten{R}$, we leverage the equivalence relation pointed out in \cite{Cook2007},
\begin{equation}\label{eq:inverse-regression-sdr}
    Y\mid\ten{X} \,{\buildrel d \over =}\, Y\mid \ten{R}(\ten{X})
        \quad\Longleftrightarrow\quad
    \ten{X}\mid(Y, \ten{R}(\ten{X})) \,{\buildrel d \over =}\,  \ten{X}\mid\ten{R}(\ten{X}).
\end{equation}
According to \eqref{eq:inverse-regression-sdr}, a \textit{sufficient statistic} $\ten{R}(\ten{X})$ for $Y$ in the inverse regression $\ten{X}\mid Y$, where $Y$ is considered as a parameter indexing the model, is also a \textit{sufficient reduction} for $\ten{X}$ in the forward regression $Y\mid\ten{X}$.

The factorization theorem is the usual tool to identify sufficient statistics and requires a distributional model. %We assume the distribution of $\ten{X}\mid Y$ belongs to the \emph{quadratic exponential family} to (a) simplify modeling and (b) keep estimation feasible. 
We assume that $\ten{X}\mid Y$ is a full rank quadratic exponential family with density
% \begin{align}
%     % f_{\mat{\eta}_y}(\ten{X}\mid Y = y)&= h(\ten{X})\exp(\t{\mat{\eta}_y}\mat{t}(\ten{X}) - b(\mat{\eta}_y)) \notag\\
%     f(\ten{X}\mid Y = y)
%     &= h(\ten{X})\exp(\t{\mat{\eta}_y}\mat{t}(\ten{X}) - b(\mat{\eta}_y)) \notag\\
%     &= h(\ten{X})\exp(\langle \mat{t}_1(\ten{X}), \mat{\eta}_{1y} \rangle + \langle \mat{t}_2(\ten{X}), \mat{\eta}_{2y} \rangle - b(\mat{\eta}_{y})) \label{eq:quad-density} %\nonumber
% \end{align}
% where $\mat{t}_1(\ten{X})=\vec \ten{X}$ and $\mat{t}_2(\ten{X})$ is linear in $\ten{X}\circ\ten{X}$.
\begin{align}
    f(\ten{X}\mid Y = y)
    &= h(\ten{X})\exp(\t{\mat{\eta}_y}\mat{t}(\ten{X}) - b(\mat{\eta}_y)) \notag \\
    &= h(\ten{X})\exp(\langle \vec(\ten{X}), \mat{\eta}_{1y} \rangle + \langle \vech(\t{(\vec\ten{X})}(\vec\ten{X})), \mat{\eta}_{2y} \rangle - b(\mat{\eta}_{y})), \label{eq:quad-density}
\end{align}
where $\mat{t}(\ten{X}) = (\vec\ten{X}, \vech(\t{(\vec\ten{X})}(\vec\ten{X})))$ with $\vech{\mat{A}}$ denoting the \emph{half-vectorization} of a matrix $\mat{A}$. The dependence of $\ten{X}$ on $Y$ is fully captured in the natural parameter $\mat{\eta}_y$.\footnote{$\mat{\eta}_y$ is a function of the response $Y$, so it is not a parameter in the formal statistical sense. We view it as a parameter in order to leverage \eqref{eq:inverse-regression-sdr} and derive a sufficient reduction from the inverse regression.}  The function $h$ is non-negative real-valued and $b$ is assumed to be at least twice continuously differentiable and strictly convex. An important feature of the \emph{quadratic exponential family} is that the distribution of its members is fully characterized by their first two moments. Distributions within the quadratic exponential family include the \emph{multi-linear normal} (\cref{sec:tensor-normal-estimation}) and \emph{multi-linear Ising model} (\cref{sec:ising_estimation}, a generalization of the (inverse) Ising model which is a multi-variate Bernoulli with up to second-order interactions) and mixtures of these two.

It is straightforward to generalize the formulation from the quadratic exponential family to incorporate higher moments. This, though, would make the number of parameters prohibitively large to estimate. This is why we opted for the quadratic exponential family in our formulation.

% The quadratic exponential family in \eqref{eq:quadratic-exp-fam} is easily generalizable to incorporate higher moments. This, though, would result in the number of parameters becoming prohibitive to estimate, which is also the reason why we opted for the quadratic exponential family in our formulation.

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{The Generalized Multi-Linear Model}\label{sec:gmlm-model}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
The dependence of $\ten{X}$ and $Y$ in model \eqref{eq:quad-density} is absorbed in $\mat{\eta}_y$, and $\mat{t}(\ten{X})$ is the minimal sufficient statistic for $\mat{\eta}_y = (\mat{\eta}_{1y}, \mat{\eta}_{2y})$. The first natural parameter component, $\mat{\eta}_{1y}$, captures the first order, and $\mat{\eta}_{2y}$, the second-order relationship of $Y$ and $\ten{X}$.

By the equivalence in \eqref{eq:inverse-regression-sdr}, in order to find the sufficient reduction $\ten{R}(\ten{X})$ we need to infer $\mat{\eta}_{1y}$, and $\mat{\eta}_{2y}$. This is reminiscent of generalized linear modeling, which we extend to a multi-linear formulation next, while deferring requirements on the parameterization to \cref{sec:manifolds}.

Suppose $\ten{F}_y$ is a known mapping of $y$ with zero expectation, $\E_Y\ten{F}_Y = 0$. We assume the dependence of $\ten{X}$ and $Y$ is reflected only in $\mat{\eta}_{1y}$ and let
\begin{equation}\label{eq:eta1-manifold}
    \mat{\eta}_{1y} = \vec{\overline{\ten{\eta}}} + \mat{B}\vec\ten{F}_y,
\end{equation}
where $\overline{\ten{\eta}}\in\mathbb{R}^{p_1\times\ldots\times p_r}$ and $\mat{B}\in\mathbb{R}^{p\times q}$. The second parameter component $\mat{\eta}_{2y} = \mat{\eta}_{2}$ is assumed to be independent of $Y$ and related to a symmetric matrix $\mat{\Omega}\in\mathbb{R}^{p\times p}$ as
\begin{equation}\label{eq:eta2-manifold}
    \mat{\eta}_{2} = c\t{\mat{D}_p}\vec\mat{\Omega},
\end{equation}
with $c\in\mathbb{R}$ a non-zero known constant determined by the distribution to ease modeling. The matrix $\mat{D}_p$ is the \emph{duplication matrix} \cite[][Ch.~11]{AbadirMagnus2005}, defined so that $\mat{D}_p\vech \mat{A} = \vec \mat{A}$ for every symmetric $p\times p$ matrix $\mat{A}$ that ensures the formulation of the density in the newly introduced parameters to have a nicer form. For example, for the multi-variate normal distribution, if the constant $c$ is chosen as $-1 / 2$, then $\mat{\Omega}$ is the inverse covariance. % That is, we assume that only $\mat{\eta}_{1y}$ depends on $Y$ through $\mat{B}$. The second parameter $\mat{\eta}_2$ captures the second-order interaction structure of $\ten{X}$, which we assume not to depend on the response $Y$.

To relate individual modes of $\ten{X}$ to the response through a multi-linear relation to $\ten{F}_y$, we assume $\ten{F}_y$ takes values in $\mathbb{R}^{q_1\times ...\times q_r}$; that is, $\ten{F}_y$ is tensor-valued. This, in turn, leads to imposing a corresponding Kronecker structure to the regression parameter $\mat{B}$ and \eqref{eq:eta1-manifold} becomes
\begin{equation}\label{eq:eta1}
    \mat{\eta}_{1y} = \vec\biggl(\overline{\ten{\eta}} + \ten{F}_y\mlm_{j = 1}^{r}\mat{\beta}_j\biggr),
\end{equation}
with $\mat{B} = \bigotimes_{j = r}^{1}\mat{\beta}_j$ and  component matrices $\mat{\beta}_j\in\mathbb{R}^{p_j\times q_j}$ for $j = 1, \ldots, r$. % are of known rank

As the bilinear form of the matrix normal requires its covariance be separable, the multi-linear structure of $\ten{X}$ also induces separability on its covariance structure (see, e.g., \cite{Hoff2011}). Therefore, we further assume that
\begin{equation}\label{eq:eta2}
    \mat{\eta}_{2} = c\t{\mat{D}_p}\vec\bigotimes_{j = r}^{1}\mat{\Omega}_j,
\end{equation}
where $\mat{\Omega}_j\in\mathbb{R}^{p_j\times p_j}$ are symmetric for $j = 1, \ldots, r$.

Writing the density in \eqref{eq:quad-density} in terms of the new parameters $\overline{\ten{\eta}}$,  $\mat{\beta}_j$, $\mat{\Omega}_j$ for $j = 1, \ldots, r$, obtains
\begin{equation}\label{eq:gmlm-density}
    f(\ten{X}\mid Y = y) % f_{\mat{\eta}_y}(\ten{X}\mid Y = y)
    = h(\ten{X})\exp\biggl(
        \biggl\langle \ten{X}, \overline{\ten{\eta}} + \ten{F}_y\mlm_{j = 1}^r\mat{\beta}_j \biggr\rangle + c\biggl\langle \ten{X}, \ten{X}\mlm_{j = 1}^r\mat{\Omega}_j \biggr\rangle - b(\mat{\eta}_y)
    \biggr),
\end{equation}
where $\mat{\eta}_y = \mat{\eta}_y(\overline{\ten{\eta}}$, $\mat{\beta}_1$, $\ldots$, $\mat{\beta}_r$, $\mat{\Omega}_1$, $\ldots$, $\mat{\Omega}_r)$ is a well defined function.  
The density of $\ten{X}$ given $Y$ in \eqref{eq:gmlm-density} is now indexed by these new parameters and sets the problem in the framework of generalized linear modeling based on a mode-wise linear relation between the predictors $\ten{X}$ and the response $Y$, which we call the \emph{Generalized Multi-Linear Model} (GMLM). Under the GMLM inverse regression model, a sufficient reduction for the forward regression of $Y$ on $\ten{X}$ is given in \cref{thm:sdr}.

\begin{theorem}[\hyperlink{proof:sdr}{SDR}]\label{thm:sdr}
    A sufficient reduction for the regression $Y\mid \ten{X}$ under the quadratic exponential family inverse regression model \eqref{eq:gmlm-density} is
    \begin{equation}\label{eq:sdr}
        \ten{R}(\ten{X}) = (\ten{X} - \E\ten{X})\mlm_{k = 1}^{r}\t{\mat{\beta}_j}.
    \end{equation}
    The reduction \eqref{eq:sdr} is minimal if all $\mat{\beta}_j$ are full rank for $j=1,\ldots,r$.
\end{theorem}

The reduction \eqref{eq:sdr} in vectorized form is $\vec\ten{R}(\ten{X})=\t{\mat{B}}\vec(\ten{X} - \E\ten{X})$. %, where $\mat{B} = \bigotimes_{k = r}^{1}\mat{\beta}_k$ with $\Span(\mat{B}) = \Span(\{\mat{\eta}_{1y} - \E_{Y}\mat{\eta}_{1Y} : y\in\mathcal{S}_Y\})$, where $\mathcal{S}_Y$ denotes the sample space of the random variable $Y$.
\cref{thm:sdr} shows that the \emph{sufficient reduction} $\ten{R}(\ten{X})$ reduces $\ten{X}$ along each mode (dimension) linearly. The graph in \cref{fig:SDRvisual} is a visual representation of the sufficient reduction for a $3$-dimensional tensor-valued predictor. We  provide the simplified version of the GMLM model and the sufficient reduction in \crefrange{ex:vector-valued}{ex:matrix-valued} for the special cases of vector- and matrix-valued predictors.

\begin{figure}[!hpt]
    \centering 
    \includegraphics[width=0.5\textwidth]{../images/reduction.pdf}
    \caption{\label{fig:SDRvisual}Visual depiction of the sufficient reduction in \cref{thm:sdr}.}
\end{figure}

% \begin{figure}[htbp]
%     \begin{minipage}{0.475\textwidth}
%         \centering 
%         \includegraphics[width=\textwidth]{../images/reduction.pdf} 
%         \caption{\label{fig:SDRvisual}Visual depiction of the sufficient reduction in \cref{thm:sdr}.}
%     \end{minipage}%
%     \hfill%
%     \begin{minipage}{0.475\textwidth}
%         \centering 
%         \includegraphics[width=\textwidth]{../plots/sim-ising-perft-m2.pdf} 
%         \caption{\label{fig:ising-m2-perft}Performance test for computing/estimating the second moment of the Ising model of dimension $p$ using either the exact method or a Monte-Carlo (MC) simulation.}
%     \end{minipage}
% \end{figure}


\begin{example}[Vector valued $\mat{x}$ ($r = 1$)]\label{ex:vector-valued}
    For a vector-valued predictor $\mat{X}\in\mathbb{R}^{p_1}$, the density \eqref{eq:gmlm-density} reduces to
    \begin{align*}
        f(\mat{x}\mid Y = y) &= h(\mat{x})\exp(\langle\mat{x}, \overline{\mat{\eta}} + \mat{\beta}_1\mat{f}_y\rangle + c\langle\mat{x}, \mat{\Omega}_1\mat{x}\rangle - b(\mat{\eta}_y)) \\
        &= h(\mat{x})\exp( \t{(\overline{\mat{\eta}} + \mat{\beta}_1\mat{f}_y)}\mat{x} + c\t{\mat{x}}\mat{\Omega}_1\mat{x} - b(\mat{\eta}_y) )
    \end{align*}
    where $\mat{f}_y\in\mathbb{R}^{q_1}$ is vector-valued as well. The sufficient reduction obtained by \cref{thm:sdr} is then $\mat{R}(\mat{x}) = \t{\mat{\beta}_1}(\mat{x} - \E\mat{X})\in\mathbb{R}^{q_1}$ and $\mat{B} = \mat{\beta}_1\in\mathbb{R}^{p_1\times q_1}$.
\end{example}

\begin{example}[Matrix-valued $\mat{X}$ ($r = 2$)]\label{ex:matrix-valued}
    Assuming $\mat{X}$ is matrix-valued, which requires $\mat{F}_Y\in\mathbb{R}^{q_1\times q_2}$ to also be matrix-valued. Then, the density \eqref{eq:gmlm-density} has the form
    \begin{align*}
        f(\mat{x}\mid Y = y)
            &= h(\mat{x})\exp(\langle\mat{x}, \overline{\mat{\eta}} + \mat{\beta}_1\mat{F}_y\t{\mat{\beta}_2}\rangle + c\langle\mat{x}, \mat{\Omega}_1\mat{x}\mat{\Omega}_2\rangle - b(\mat{\eta}_y)) \\
            &= h(\mat{x})\exp(\tr((\overline{\mat{\eta}} + \mat{\beta}_1\mat{F}_y\t{\mat{\beta}_2})\t{\mat{x}}) + c \tr(\mat{\Omega}_1\mat{x}\mat{\Omega}_2\t{\mat{x}}) - b(\mat{\eta}_y(\mat{\theta})))
    \end{align*}
    where $\tr(\mat{A})$ is the trace of a square matrix $\mat{A}$. By \Cref{thm:sdr}, the sufficient reduction is $\mat{R}(\mat{X}) = \t{\mat{\beta}_1}(\mat{X} - \E\mat{X})\mat{\beta}_2\in\mathbb{R}^{q_1\times q_2}$, or in vector form, $\vec\mat{R}(\mat{X}) = \t{\mat{B}}\vec(\mat{X} - \E\mat{X})$ with $\mat{B} = \mat{\beta}_2\otimes\mat{\beta}_1\in\mathbb{R}^{p_1 p_2\times q_1 q_2}$.
\end{example}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Manifolds and Parameter Spaces}\label{sec:manifolds}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
%We start by pinpointing an important issue in the GMLM formulation: The parameters are non-identifiable. We resolve this by connecting the GMLM parameters to identifiable ones living in a space with special structure. This leads to imposing conditions to the GMLM parameters that result from the relation between the new identifiable parameter and the original GMLM parameters.

\cref{thm:sdr} finds the sufficient reduction for the regression of $Y$ on $\ten{X}$ in the population under the inverse GMLM \eqref{eq:gmlm-density}. In practice we need to estimate the mode-wise reduction matrices $\mat{\beta}_j$  in the GMLM \eqref{eq:gmlm-density}. As we operate within the framework of the exponential family, we opt for maximum likelihood estimation (MLE). In a classic generalized linear model setting, this is straightforward and yields well defined MLEs. In our setting, though, a major problem is due to the fact that our GMLM parameterization in \eqref{eq:gmlm-density} is \emph{not} identifiable. This is a direct consequence of the identity $\mat{\beta}_2\otimes\mat{\beta}_1 = (a\mat{\beta}_2)\otimes (a^{-1}\mat{\beta}_1)$ for any $a\neq 0$ (the same holds for the $\mat{\Omega}_j$s) so that different parameterizations of \eqref{eq:gmlm-density} describe the same density. In other words, we do \emph{not} have a one-to-one relation between the parameters and the GMLM and consistency cannot be established. Without consistency, derivation of the asymptotic distribution for the maximum likelihood estimator becomes infeasible.

To resolve this issue, we need to disambiguate the GMLM parameters to reestablish a one-to-one relation to the model. At the same time, we want to keep the mode-wise parameters $\mat{\beta}_j$ in \eqref{eq:gmlm-density} as those are the mode-wise reduction matrices needed by \cref{thm:sdr}. Using the mode-wise GMLM parameters has further advantages: the total number of parameters to be estimated is much smaller in many settings. Specifically, in the case of the multi-linear normal, a very efficient estimation algorithm is applicable. Moreover, the required number of observations for a reliable estimate is very small, potentially even smaller than any of the axis-dimensions $p_j$. We also gain significant estimation accuracy. % if the modeling assumptions are valid.

In the derivation of the GMLM we first introduced the parameters $\overline{\eta}$ and $\mat{B}$,  which models a linear relation between $\ten{X}$ and $Y$ through $\mat{\eta}_{1y}$ in \eqref{eq:eta1-manifold}, and  the symmetric matrix $\mat{\Omega}$ to replace $\mat{\eta}_2$ in \eqref{eq:eta2-manifold}. Then, we modeled $\mat{B}$ using the mode-wise component matrices $\mat{\beta}_j$ which impose a non-linear constraint on $\mat{B} = \bigotimes_{j = r}^1\mat{\beta}_j$. Similarly, the introduction of the $\mat{\Omega}_j$'s in $\mat{\Omega} = \bigotimes_{j = r}^1 \mat{\Omega}_j$ constrains nonlinearly $\mat{\Omega}$. Both unconstrained $\mat{B}$ and $\vech(\mat{\Omega})$ are identifiable: The GMLM density \eqref{eq:gmlm-density} corresponds to one and only one  $\mat{B}$ and $\vech(\mat{\Omega})$. Additionally, given any $\mat{\beta}_j$'s and $\mat{\Omega}_j$'s, then $\mat{B}$ and $\vech(\mat{\Omega})$ are uniquely determined. Based on these observations we derive the asymptotic behavior of the parameters $\mat{B}$ and $\vech{\mat{\Omega}}$ while operating with their components $\mat{\beta}_j$ and $\mat{\Omega}_j$. As a result we obtain a parameter space $\Theta$ with a non-linear constraint.

% Any estimation of the sufficient reduction requires application of some optimality criterion. As we operate within the framework of the exponential family, we opted for maximum likelihood estimation (MLE). For the unconstrained problem, where the parameters are simply $\mat{B}$ and $\mat{\Omega}$ in \eqref{eq:eta1-manifold}, maximizing the likelihood of $\ten{X} \mid Y$ is straightforward and yields well-defined MLEs of both parameters. Our setting, though, requires the constrained optimization of the $\ten{X} \mid Y$ likelihood subject to  $\mat{B} = \bigotimes_{j = r}^{1}\mat{\beta}_j$ and $\mat{\Omega}=\bigkron_{j = r}^{1}\mat{\Omega}_j$. \Cref{thm:kron-manifolds,thm:param-manifold} provide the setting for which the MLE of the constrained parameter $\mat{\theta}$ is well-defined, which in turn leads to the derivation of its asymptotic normality.

% The main problem in obtaining asymptotic results for the MLE of the constrained parameter $\mat{\theta} = (\overline{\ten{\eta}}, \vec\mat{B}, \vech\mat{\Omega})$ stems from the nature of the constraint. We assumed that $\mat{B} = \bigkron_{k = r}^{1}\mat{\beta}_k$, where the parameter $\mat{B}$ is identifiable. This means that different values of $\mat{B}$ lead to different densities $f_{\mat{\theta}}(\ten{X}\mid Y = y)$, a basic property needed to ensure consistency of parameter estimates, which in turn is needed for asymptotic normality. On the other hand, the components $\mat{\beta}_j$, $j = 1, \ldots, r$, are \emph{not} identifiable, which is a direct consequence of the equality $\mat{\beta}_2\otimes\mat{\beta}_1 = (c\mat{\beta}_2)\otimes (c^{-1}\mat{\beta}_1)$ for every $c\neq 0$. This is the reason we considered  $\Theta$ as a constrained parameter space instead of parameterizing the densities of $\ten{X}\mid Y$ with $\mat{\beta}_1, \ldots, \mat{\beta}_r$. The same is true for $\mat{\Omega} = \bigkron_{k = r}^{1}\mat{\Omega}_k$.

Except for identifiable parameters, asymptotic normality (see \cref{thm:asymptotic-normality-gmlm} in  \cref{sec:statprop})  requires differentiation. Therefore, the space itself must admit the definition of differentiation, which is usually a vector space. This is too strong an assumption for our purposes. To weaken the vector space assumption, we consider \emph{smooth manifolds}. These are spaces that look like Euclidean spaces locally and allow the notion of differentiation. The more general \emph{topological} manifolds are too weak for differentiation. A smooth manifold only allows only for first derivatives. Without going into details, the solution is a \emph{Riemannian manifold} \cite[]{Lee2012,Lee2018,AbsilEtAl2007}. Similar to an abstract \emph{smooth manifold}, Riemannian manifolds are detached from our usual intuition as well as are complicated to handle. This is where an \emph{embedded (sub)manifold} comes to the rescue. Simply speaking, an embedded manifold is a manifold that is a subset of a manifold from which it inherits its properties. If a manifold is embedded in a Euclidean space, almost all the complications of abstract manifold theory simplify drastically. Moreover, since an Euclidean space is itself a Riemannian manifold, we inherit the means for higher derivatives. Finally, a smooth embedded submanifold structure for the parameter space maintains consistency with existing approaches and results for parameter sets with linear subspace structure. These reasons justify the constraint that the parameter space $\Theta$ be a \emph{smooth embedded manifold} in a Euclidean space. % an open subset $\Xi$ of 

We now define a \emph{smooth manifold} embedded in $\mathbb{R}^p$, avoiding unnecessary detours into the more general theory (see \cite{Kaltenbaeck2021}).

\begin{definition}[Embedded smooth manifold]\label{def:manifold}
    A set $\manifold{A}\subseteq\mathbb{R}^p$ is an \emph{embedded smooth manifold} of dimension $d$ if for every $\mat{x}\in\manifold{A}$ there exists a smooth\footnote{Here \emph{smooth} means infinitely differentiable or $C^{\infty}$.} bi-continuous map $\varphi:U\cap\manifold{A}\to V$, called a \emph{chart}, with $\mat{x}\in U\subseteq\mathbb{R}^p$ open and $V\subseteq\mathbb{R}^d$ open.
\end{definition}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Kronecker Product Manifolds}\label{sec:kron-manifolds}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

As a basis to ensure that the constrained parameter space $\Theta$ is a manifold, without having to check every case separately, we provide two simple criteria that can be used to construct manifolds obeying the Kronecker product constraint. We need the concept of a \emph{spherical} set, which is a set $\manifold{A}$, on which the Frobenius norm, $\|\,.\,\|_F:\manifold{A}\to\mathbb{R}$, is constant. Also, we call a scale invariant set $\manifold{A}$ a \emph{cone}; that is, $\manifold{A} = \{ c \mat{A} : \mat{A}\in\manifold{A} \}$ for all $c > 0$.

% As a basis to ensure that the constrained parameter space $\Theta$ is a manifold, which is a requirement of \cref{thm:param-manifold}, we need \cref{thm:kron-manifolds}. Therefore, we need the notion of a \emph{spherical} set, which is a set $\manifold{A}$, on which the Frobenius norm, $\|\,.\,\|_F:\manifold{A}\to\mathbb{R}$, is constant.  Furthermore, we call a scale invariant set $\manifold{A}$ a \emph{cone}, that is $\manifold{A} = \{ c \mat{A} : \mat{A}\in\manifold{A} \}$ for all $c > 0$.

\begin{theorem}[\hyperlink{proof:kron-manifolds}{Kronecker Product Manifolds}]\label{thm:kron-manifolds}
    Let $\manifold{A}\subseteq\mathbb{R}^{p_1\times q_1}\backslash\{\mat{0}\}, \manifold{B}\subseteq\mathbb{R}^{p_2\times q_2}\backslash\{\mat{0}\}$ be smooth embedded submanifolds. Assume one of the following conditions holds:
    \begin{itemize}
        \item[-] ``sphere condition'':
            At least one of $\manifold{A}$ or $\manifold{B}$ is \emph{spherical} and let $d = \dim\manifold{A} + \dim\manifold{B}$.
        \item[-] ``cone condition'':
            Both $\manifold{A}$ and $\manifold{B}$ are \emph{cones} and let $d = \dim\manifold{A} + \dim\manifold{B} - 1$.
    \end{itemize}
    Then, $\{ \mat{A}\otimes \mat{B} : \mat{A}\in\manifold{A}, \mat{B}\in\manifold{B} \}\subset\mathbb{R}^{p_1 p_2\times q_1 q_2}$ is a smooth embedded $d$-manifold.
\end{theorem}

With \cref{thm:kron-manifolds} we can obtain sufficient conditions for the construction of a constrained parameter manifold.

\begin{theorem}[\hyperlink{proof:param-manifold}{Parameter Manifolds}]\label{thm:param-manifold}
    Let
    \begin{displaymath}
        \manifold{K}_{\mat{B}} = \Bigl\{ \bigkron_{k = r}^{1}\mat{\beta}_k : \mat{\beta}_k\in\manifold{B}_k \Bigr\}
        \quad\text{and}\quad
        \manifold{K}_{\mat{\Omega}} = \Bigl\{ \bigkron_{k = r}^{1}\mat{\Omega}_k : \mat{\Omega}_k\in\manifold{O}_k \Bigr\}
    \end{displaymath}
    where $\manifold{B}_k\subseteq\mathbb{R}^{p_k\times q_k}\backslash\{\mat{0}\}$ and $\manifold{O}_k\subseteq\SymMat{p_k}\backslash\{\mat{0}\}$ are smooth embedded manifolds that are either spheres or cones, for $k = 1, ..., r$.  Then, the \emph{constrained parameter space} \[ \Theta = \mathbb{R}^p \times \manifold{K}_{\mat{B}}\times\vech(\manifold{K}_{\mat{\Omega}})\subset\mathbb{R}^{p(p + 2 q + 3) / 2}\] 
    as well as $\manifold{K}_{\mat{B}}$, $\manifold{K}_{\mat{\Omega}}$ and $\vech{\manifold{K}_{\mat{\Omega}}}$ are smooth embedded manifolds, where $\vech(\manifold{K}_{\mat{\Omega}}) = \{ \vech{\mat{\Omega}} : \mat{\Omega}\in\manifold{K}_{\mat{\Omega}} \}$.
\end{theorem}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Matrix Manifolds}\label{sec:matrix-manifolds}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
A powerful feature of \cref{thm:param-manifold} is the modeling flexibility it provides. For example, we can perform low-rank regression. Or, we may constrain two-way interactions between direct axis neighbors by using band matrices for the $\mat{\Omega}_k$'s, among others.

This flexibility derives from many different matrix manifolds that can be used as building blocks $\manifold{B}_k$ and $\manifold{O}_k$ of the parameter space $\Theta$ in \cref{thm:param-manifold}. A list of possible choices, among others, is given in \cref{tab:matrix-manifolds}. As long as parameters in $\Theta$ are a valid parameterization of a density (or probability mass function) of \eqref{eq:quad-density} subject to \eqref{eq:eta1-manifold} and \eqref{eq:eta2-manifold}, one may choose any of the manifolds listed in \cref{tab:matrix-manifolds} which are either cones or spherical. We also include an example which is neither a sphere nor a cone. They may also be valid building blocks but require more work as they are not directly leading to a parameter manifold by \cref{thm:param-manifold}. If the resulting parameter space $\Theta$ is an embedded manifold, the asymptotic theory of \cref{sec:statprop} is applicable.

\begin{table}
    \caption{\label{tab:matrix-manifolds}Examples of embedded matrix manifolds. ``Symbol'' a (more or less) common notation for the matrix manifold, if at all. ``C'' stands for \emph{cone}, meaning it is scale invariant. ``S'' means \emph{spherical}, that is, constant Frobenius norm.}
    \begin{tabular}{l l c c c}
        \hline
        Symbol & Description & C & S & Dimension\\
        \hline
        $\mathbb{R}^{p\times q}$ & All matrices of dimension $p\times q$ &
            \checkmark & \xmark     & $p q$ \\
            $\mathbb{R}_{*}^{p\times q}$ & Full rank $p\times q$ matrices &
            \checkmark & \xmark     & $p q$                  \\
        $\Stiefel{p}{q}$ & \emph{Stiefel Manifold}, $\{ \mat{U}\in\mathbb{R}^{p\times q} : \t{\mat{U}}\mat{U} = \mat{I}_q \}$ for $q\leq p$ &
            \xmark     & \checkmark & $p q - q (q + 1) / 2$  \\
        $\mathcal{S}^{p - 1}$ & Unit sphere in $\mathbb{R}^p$, special case $\Stiefel{p}{1}$ &
            \xmark     & \checkmark & $p - 1$                \\
        $\OrthogonalGrp{p}$ & Orthogonal Group, special case $\Stiefel{p}{p}$ &
            \xmark     & \checkmark & $p (p - 1) / 2$        \\
        $\SpecialOrthogonalGrp{p}$ & Special Orthogonal Group $\{ \mat{U}\in U(p) : \det{\mat{U}} = 1 \}$ &
            \xmark     & \checkmark & $p (p - 1) / 2$        \\
        $\mathbb{R}_{r}^{p\times q}$ & Matrices of known rank $r > 0$, generalizes $\StiefelNonCompact{p}{q}$ &
            \checkmark & \xmark     & $r(p + q - r)$         \\
        $\SymMat{p}$ & Symmetric matrices &
            \checkmark & \xmark     & $p (p + 1) / 2$    \\
        $\SymPosDefMat{p}$ & Symmetric Positive Definite matrices &
            \checkmark & \xmark     & $p (p + 1) / 2$    \\
        & Scaled Identity $\{ a\mat{I}_p : a\in\mathbb{R}_{+} \}$ &
            \checkmark & \xmark     & $1$    \\
        & Symmetric $r$-band matrices (includes diagonal) &
            \checkmark & \xmark     & $(2 p - r) (r + 1) / 2$    \\
        & Auto correlation $\{ \mat{A}\in\mathbb{R}^{p\times p} : \mat{A}_{i j} = \rho^{|i - j|}, \rho\in(0, 1) \}$ &
            \xmark     & \xmark     & $1$ \\
        \hline
    \end{tabular}
\end{table}

\begin{remark}
    The \emph{Grassmann Manifold} of $q$ dimensional subspaces in $\mathbb{R}^p$ is not listed in \cref{tab:matrix-manifolds} since it is not embedded in $\mathbb{R}^{p \times q}$.
\end{remark}



%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Maximum Likelihood Estimation}\label{sec:ml-estimation}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
Suppose $(\ten{X}_i, Y_i)$ are independently and identically distributed with joint cdf $F(\ten{X}, Y)$, for $i = 1, \ldots, n$. The empirical log-likelihood function of \eqref{eq:gmlm-density}, ignoring terms not depending on the parameters, is
\begin{equation}\label{eq:log-likelihood}
    l_n(\mat{\theta}) = \frac{1}{n}\sum_{i = 1}^n \biggl(\Bigl\langle\ten{X}_i, \overline{\ten{\eta}} + \ten{F}_{y_i}\mlm_{k = 1}^{r}\mat{\beta}_k \Bigr\rangle + c\Bigl\langle\ten{X}_i, \ten{X}_i\mlm_{k = 1}^{r}\mat{\Omega}_k \Bigr\rangle - b(\mat{\eta}_{y_i})\biggr).
\end{equation}
The maximum likelihood estimate of $\mat{\theta}_0=(\vec\overline{\ten{\eta}}_0, \vec\mat{B}_0, \vech\mat{\Omega}_0)$ is the solution to the optimization problem
\begin{equation}\label{eq:mle}
    \hat{\mat{\theta}}_n = \argmax_{\mat{\theta}\in\Theta}l_n(\mat{\theta})
\end{equation}
with $\hat{\mat{\theta}}_n = (\vec\widehat{\overline{\ten{\eta}}}, \vec\widehat{\mat{B}}, \vech\widehat{\mat{\Omega}})$ where $\widehat{\mat{B}} = \bigkron_{k = r}^{1}\widehat{\mat{\beta}}_k$ and $\widehat{\mat{\Omega}} = \bigkron_{k = r}^{1}\widehat{\mat{\Omega}}_k$.

In a classical \emph{generalized linear model} (GLM), the link function connecting the natural parameters to the expectation of the sufficient statistic $\mat{\eta}_y = \mat{g}(\E[\mat{t}(\ten{X}) \mid Y = y])$ is invertible. Such a link may not exist in our setting, but for our purpose it is sufficient to know the conditional first $\E_{\mat{\theta}}[\ten{X} \mid Y = y]$  and second $\E_{\mat{\theta}}[\ten{X}\circ\ten{X} \mid Y = y]$ moments, given a parameterization $\mat{\theta}$ of \eqref{eq:gmlm-density}.

Gradient descent is a powerful and widely used optimization algorithm  to compute MLEs. %A straightforward general method for parameter estimation is \emph{gradient descent}. 
%To apply gradient-based optimization, 
We compute the gradients of $l_n$ in \cref{thm:grad}.

\begin{theorem}[\hyperlink{proof:grad}{Likelihood Gradient}]\label{thm:grad}
    Suppose $(\ten{X}_i, y_i), i = 1, ..., n$, are i.i.d. with conditional log-likelihood of the form \eqref{eq:log-likelihood}, where $\mat{\theta}$ denotes the collection of all GMLM parameters $\overline{\ten{\eta}}$, ${\mat{B}} = \bigkron_{k = r}^{1}{\mat{\beta}}_k$ and ${\mat{\Omega}} = \bigkron_{k = r}^{1}{\mat{\Omega}}_k$ for $k = 1, ..., r$. Then, the partial gradients with respect to $\overline{\ten{\eta}}, \mat{\beta}_1, \ldots, \mat{\beta}_r, \mat{\Omega}_1, \ldots, \mat{\Omega}_r$ are given by
    \begin{align*}
        \nabla_{\overline{\ten{\eta}}}l_n &\equiv \frac{1}{n}\sum_{i = 1}^n (\ten{X}_i - \E_{\mat{\theta}}[\ten{X} \mid Y = y_i]), \\
        \nabla_{\mat{\beta}_j}l_n &\equiv \frac{1}{n}\sum_{i = 1}^n (\ten{X}_i - \E_{\mat{\theta}}[\ten{X} \mid Y = y_i])_{(j)}\t{\Big(\ten{F}_{y_i}\mlm_{\substack{k = 1\\k\neq j}}^r\mat{\beta}_k\Big)_{(j)}}, \\
        \nabla_{\mat{\Omega}_j}l_n &\equiv \frac{c}{n}\sum_{i = 1}^n (\ten{X}_i\otimes\ten{X}_i - \E_{\mat{\theta}}[\ten{X}\otimes\ten{X} \mid Y = y_i])\mlm_{\substack{k = 1\\k\neq j}}^r\t{(\vec{\mat{\Omega}_k})},
    \end{align*}
    so that  $\nabla l_n = (\nabla_{\overline{\ten{\eta}}}l_n, \nabla_{\mat{\beta}_1}l_n, \ldots, \nabla_{\mat{\beta}_r}l_n, \nabla_{\mat{\Omega}_1}l_n, \ldots, \nabla_{\mat{\Omega}_r}l_n)$.
\end{theorem}

The partial gradients $\nabla_{\mat{\Omega}_j}l_n$ contain the \emph{Kronecker} product $\ten{X}\otimes\ten{X}$ of tensor-valued predictors $\ten{X}$ instead of the outer product, which differ only in the order of their elements and shape.

Although  any GMLM can be fitted via gradient descent using \cref{thm:grad}, this may be computationally inefficient. In the special case of multi-linear normal predictors, an iterative cyclic updating scheme that converges fast, is stable, and does not require hyperparameters is derived in \cref{sec:tensor-normal-estimation}. On the other hand, the Ising model does not allow such a scheme, requiring a gradient-based method.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Multi-Linear Normal}\label{sec:tensor-normal-estimation}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
The \emph{multi-linear normal} is the extension of the matrix normal to tensor-valued random variables and a member of the quadratic exponential family \eqref{eq:quad-density} under \eqref{eq:eta2}. \cite{Dawid1981} and \cite{Arnold1981} introduced the term matrix normal and, in particular, \cite{Arnold1981} provided several theoretical results, such as its density, moments and conditional distributions of its components. The matrix normal distribution is  a bilinear normal distribution; a distribution of a two-way (two-component) array, each component representing a vector of observations \cite[]{OhlsonEtAl2013}. \cite{KolloVonRosen2005,Hoff2011,OhlsonEtAl2013} presented the extension of the bilinear to the multi-linear normal distribution, using a parallel extension of bilinear matrices to multi-linear tensors \cite[]{Comon2009}.

The defining feature of the matrix normal distribution, and its multi-linear extension, is the Kronecker product structure of its covariance. This formulation, where the vectorized variables are normal with multiway covariance structure modeled as a Kronecker product of matrices of much lower dimension, aims to overcome the significant modeling and computational challenges arising from the high computational complexity of manipulating tensor representations \cite[see, e.g.,][]{HillarLim2013,WangEtAl2022}.

Suppose the conditional distribution of tensor $\ten{X}$ given $Y$ is multi-linear normal  with mean $\ten{\mu}_y$ and covariance $\mat{\Sigma} = \bigkron_{k = r}^{1}\mat{\Sigma}_k$. We assume the distribution is non-degenerate which means that the covariances $\mat{\Sigma}_k$ are symmetric positive definite matrices. Its density is
\begin{displaymath}
    f_{\mat{\theta}}(\ten{X}\mid Y = y) = (2\pi)^{-p / 2}\prod_{k = 1}^{r}\det(\mat{\Sigma}_k)^{-p / 2 p_k}\exp\biggl( -\frac{1}{2}\biggl\langle\ten{X} - \ten{\mu}_y, (\ten{X} - \ten{\mu}_y)\mlm_{k = 1}^{r}\mat{\Sigma}_k^{-1} \biggr\rangle \biggr).
\end{displaymath}
For the sake of simplicity and w.l.o.g., we assume $\ten{X}$ has 0 marginal expectation; i.e., $\E\ten{X} = 0$. Rewriting the multi-linear normal density in the form of the GMLM density \eqref{eq:gmlm-density} with the scaling constant $c = -1/2$ yields the parameter relations
\begin{equation}\label{eq:tnormal_cond_params}
    \ten{\mu}_y = \ten{F}_y\mlm_{k = 1}^{r}\mat{\Omega}_k^{-1}\mat{\beta}_k, \qquad
    \mat{\Omega}_k = \mat{\Sigma}_k^{-1}.
\end{equation}
Here we used that $\overline{\ten{\eta}} = 0$ due to $0 = \E\ten{X} = \E\E[\ten{X}\mid Y] = \E\ten{\mu}_Y$ in combination with $\E\ten{F}_Y = 0$. Additionally, the positive definiteness of the $\mat{\Sigma}_k$'s renders all the $\mat{\Omega}_k$'s symmetric positive definite. We obtain $\E_{\mat{\theta}}[\ten{X}\mid Y = y] = \ten{\mu}_y$, and 
\begin{displaymath}
    % \ten{G}_2(\mat{\eta}_y) = 
    \E_{\mat{\theta}}[\ten{X}\circ\ten{X}\mid Y = y] \equiv \bigkron_{k = r}^1\mat{\Sigma}_k + (\vec{\ten{\mu}}_y)\t{(\vec{\ten{\mu}}_y)}.
\end{displaymath}
In practice, we assume we have a random sample of $n$ observations $(\ten{X}_i, \ten{F}_{y_i})$ from the joint distribution. We start the estimation process by demeaning the data. Then, only the reduction matrices $\mat{\beta}_k$ and the scatter matrices $\mat{\Omega}_k$ need to be estimated. To solve the optimization problem \eqref{eq:mle}, with $\overline{\ten{\eta}} = 0$, we initialize the parameters using a simple heuristic approach. First, we compute moment based mode-wise marginal covariance estimates $\widehat{\mat{\Sigma}}_k(\ten{X})= \sum_{i = 1}^{n} (\ten{X}_i)_{(k)}\t{(\ten{X}_i)_{(k)}}/n$ and $\widehat{\mat{\Sigma}}_k(\ten{F}_Y)= \sum_{i = 1}^{n} (\ten{F}_{y_i})_{(k)}\t{(\ten{F}_{y_i})_{(k)}}/n$. % as
%\begin{displaymath}
%\widehat{\mat{\Sigma}}_k(\ten{X})   = \frac{1}{n}\sum_{i = 1}^{n} (\ten{X}_i)_{(k)}\t{(\ten{X}_i)_{(k)}}, \qquad
%\widehat{\mat{\Sigma}}_k(\ten{F}_Y) = \frac{1}{n}\sum_{i = 1}^{n} (\ten{F}_{y_i})_{(k)}\t{(\ten{F}_{y_i})_{(k)}}.
%\end{displaymath}
Then, for every mode $k = 1, \ldots, r$, we compute the  eigenvectors $\mat{v}_j(\widehat{\mat{\Sigma}}_k(\ten{X}))$, $\mat{v}_j(\widehat{\mat{\Sigma}}_k(\ten{F}_Y))$ that correspond to the leading eigenvalues $\lambda_j(\widehat{\mat{\Sigma}}_k(\ten{X}))$, $\lambda_j(\widehat{\mat{\Sigma}}_k(\ten{X}))$ of the marginal covariance estimates, for $j = 1, \ldots, q_k$. We set 
\begin{align*}
    \mat{U}_k &= (\mat{v}_1(\widehat{\mat{\Sigma}}_1(\ten{X})), \ldots, \mat{v}_{q_k}(\widehat{\mat{\Sigma}}_{q_k}(\ten{X}))), \,  \mat{V}_k = (\mat{v}_1(\widehat{\mat{\Sigma}}_1(\ten{F}_Y), \ldots, \mat{v}_{q_k}(\widehat{\mat{\Sigma}}_{q_k}(\ten{F}_Y))\\
    \mat{D}_k &= \diag(\mat{v}_1(\widehat{\mat{\Sigma}}_1(\ten{X}))\mat{v}_1(\widehat{\mat{\Sigma}}_1(\ten{F}_{Y})), \ldots, \mat{v}_{q_k}(\widehat{\mat{\Sigma}}_{q_k}(\ten{X}))\mat{v}_{q_k}(\widehat{\mat{\Sigma}}_k(\ten{F}_{Y}))). 
\end{align*}
The initial value of $\mat{\beta}_k$  is
%\begin{displaymath}
 $   \hat{\mat{\beta}}_k^{(0)} = \mat{U}_k\sqrt{\mat{D}_k}\t{\mat{V}_k},$
%\end{displaymath}
and the initial value of $\mat{\Omega}_k$ is  set to the identity $\mat{\Omega}_k^{(0)} = \mat{I}_{p_k}$, for $k=1,\ldots,r$.
Given  $\hat{\mat{\beta}}_1, \ldots, \hat{\mat{\beta}}_r, \hat{\mat{\Omega}}_1, \ldots, \hat{\mat{\Omega}}_r$, we take the gradient $\nabla_{\mat{\beta}_j}l_n$ of the multi-linear normal log-likelihood $l_n$ in \eqref{eq:log-likelihood}  applying  \cref{thm:grad} and keep all other parameters except $\mat{\beta}_j$ fixed. Then, $\nabla_{\mat{\beta}_j}l_n = 0$ has the closed form solution
\begin{equation}\label{eq:tensor_normal_beta_solution}
    \t{\mat{\beta}_j} = \biggl(
        \sum_{i = 1}^{n}
        \Bigl( \ten{F}_{y_i}\mlm_{k \neq j}\hat{\mat{\Omega}}_k^{-1}\hat{\mat{\beta}}_k \Bigr)_{(j)}
        \t{\Bigl( \ten{F}_{y_i}\mlm_{k \neq j}\hat{\mat{\beta}}_k \Bigr)_{(j)}}
    \biggr)^{-1}
    \biggl(
        \sum_{i = 1}^{n}
        \Bigl( \ten{F}_{y_i}\mlm_{k \neq j}\hat{\mat{\beta}}_k \Bigr)_{(j)}
        \t{(\ten{X}_{i})_{(j)}}
    \biggr)
        \hat{\mat{\Omega}}_j.
\end{equation}
Equating the partial gradient of the $j$th scatter matrix $\mat{\Omega}_j$ in \cref{thm:grad} to zero ($\nabla_{\mat{\Omega}_j}l_n = 0$) gives a quadratic matrix equation due to the dependence of $\ten{\mu}_y$ on $\mat{\Omega}_j$. In practice though, it is faster, more stable,  and equally accurate to use mode-wise covariance estimates via the residuals
\begin{displaymath}
    \hat{\ten{R}}_i = \ten{X}_i - \hat{\ten{\mu}}_{y_i} = \ten{X}_i - \ten{F}_{y_i}\mlm_{k = 1}^{r}\hat{\mat{\Omega}}_k^{-1}\hat{\mat{\beta}}_k.
\end{displaymath} 
The estimates are computed via 
%\begin{displaymath}
 $   \tilde{\mat{\Sigma}}_j = \sum_{i = 1}^n (\hat{\ten{R}}_i)_{(j)} \t{(\hat{\ten{R}}_i)_{(j)}},$
%\end{displaymath}
where $\tilde{s}\tilde{\mat{\Sigma}}_j = \hat{\mat{\Omega}}_j^{-1}$. To decide on the scaling factor $\tilde{s}$  we use that the mean squared error has to be equal to the trace of the covariance estimate,
\begin{displaymath}
    \frac{1}{n}\sum_{i = 1}^n \langle \hat{\ten{R}}_i, \hat{\ten{R}}_i \rangle = \tr\bigkron_{k = r}^{1}\hat{\mat{\Omega}}_k^{-1} = \prod_{k = 1}^{r}\tr{\hat{\mat{\Omega}}_k^{-1}} = \tilde{s}^r\prod_{k = 1}^{r}\tr{\tilde{\mat{\Sigma}}_k},
\end{displaymath}
so that
\begin{displaymath}
    \tilde{s} = \biggl(\Bigl(\prod_{k = 1}^{r}\tr{\tilde{\mat{\Sigma}}_k}\Bigr)^{-1}\frac{1}{n}\sum_{i = 1}^n \langle \hat{\ten{R}}_i, \hat{\ten{R}}_i \rangle\biggr)^{1 / r}
\end{displaymath}
resulting in the estimates $\hat{\mat{\Omega}}_j = (\tilde{s}\tilde{\mat{\Sigma}}_j)^{-1}$.
Estimation is performed by updating  the estimates $\hat{\mat{\beta}}_j$ via \eqref{eq:tensor_normal_beta_solution} for $j = 1, \ldots, r$,  and then recompute the $\hat{\mat{\Omega}}_j$ estimates simultaneously keeping the $\hat{\mat{\beta}}_j$s fixed. This procedure is repeated until convergence.

A technical detail for numerical stability is to ensure that the scaled values $\tilde{s}\tilde{\mat{\Sigma}}_j$, assumed to be symmetric and positive definite, are well conditioned. Thus, we estimate the condition number of $\tilde{s}\tilde{\mat{\Sigma}}_j$ before computing the inverse. In case of ill-conditioning, we use the regularized $\hat{\mat{\Omega}}_j = (\tilde{s}\tilde{\mat{\Sigma}}_j + 0.2 \lambda_{1}(\tilde{s}\tilde{\mat{\Sigma}}_j)\mat{I}_{p_j})^{-1}$ instead, where $\lambda_{1}(\tilde{s}\tilde{\mat{\Sigma}}_j)$ is the first (maximum) eigenvalue. Experiments showed that this regularization is usually only required in the first few iterations.

If the parameter space follows a more general setting as in \cref{thm:param-manifold}, updating may produce estimates outside the parameter space. A simple and efficient method is to project every updated estimate onto the corresponding manifold.

A standard algorithm to calculate the MLE of a Kronecker product is block-coordinate descent,  proposed independently by \cite{MardiaGoodall1993} and \cite{Dutilleul1999}. It was later called ``flip-flop'' algorithm by \cite{LuZimmerman2005} for the computation of the maximum likelihood estimators of the components of a separable covariance matrix. \cite{ManceurDutilleul2013} extended the ``flip-flop'' algorithm for the computation of the MLE of the separable covariance structure of a 3-way and 4-way normal distribution and obtained a lower bound for the sample size required for its existence. The same issue was also studied by \cite{DrtonEtAl2020} in the case of a two-way array (matrix). Our algorithm uses a similar ``flip-flop'' approach by iteratively updating the $\mat{\beta}_k$'s and $\mat{\Omega}_k$'s.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Multi-Linear Ising Model}\label{sec:ising_estimation}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
The Ising\footnote{Also known as the \emph{Lenz-Ising} model as the physical assumptions of the model where developed by both Lenz and Ising \cite[]{Niss2005} where Ising gave a closed form solution for the 1D lattice \cite[]{Ising1925}.} model \cite[]{Lenz1920,Ising1925,Niss2005} is a mathematical model originating in statistical physics to study ferromagnetism in a thermodynamic setting. It describes magnetic dipoles (atomic ``spins'' with values $\pm 1$) under an external magnetic field (first moments) while allowing two-way interactions (second moments) between direct neighbors on a lattice, a discrete grid. %The Ising problem, as known in statistical physics, is to compute observables such as the magnetizations and correlations under the Boltzmann distribution\footnote{The Boltzmann distribution is a probability distribution over the states of a physical system in thermal equilibrium (constant temperature) that assigns higher probabilities to states with lower energy.} while the interaction structure and the magnetic fields are given. The ``reverse'' problem, where the couplings and fields are unknown and to be determined from observations of the spins, as in statistical inference, is known as the \emph{inverse Ising problem} \cite[]{NguyenEtAl2017}. 
%From this point of view, 
The Ising model is a member of the discrete quadratic exponential family \cite[]{CoxWermuth1994,JohnsonEtAl1997} for multivariate binary outcomes where the interaction structure (non-zero correlations) is determined by the lattice. 
%Generally, neither the values of couplings nor the interaction structure are known.
%In consequence, the Ising model is mostly used to model multivariate binary data in statistics. The states are ${0, 1}$ instead of $\pm 1$, and full interaction structure. 
%It is related to a multitude of other models, among which the most prominent are: \emph{Graphical Models} and \emph{Markov Random Fields} to describe conditional dependence \cite[]{Lauritzen1996,WainwrightJordan2008,LauritzenRichardson2002}, \emph{Potts models} \cite[]{Besag1974,ChakrabortyEtAl2022} which generalize the Ising model to multiple states, the \emph{multivariate Bernoulli distribution} \cite[]{Whittaker1990,JohnsonEtAl1997,DaiDingWahba2013} that also accommodates higher-order interactions (three-way and higher), \emph{(restricted) Botlzmann machines} \cite[]{Smolensky1986,Hinton2002,FischerIgel2012} that introduce additional hidden variables for learning binary distributions. Most of these models can be used both in supervised and unsupervised settings. 
%Applications of the Ising model (and variations thereof) range from modeling neural firing patterns \cite[]{SchneidmanEtAl2006}, gene expression data analysis \cite[]{LezonEtAl2006}, and modeling financial markets \cite[]{Bury2013}. See also \cite{NguyenEtAl2017}.
%The $r$-tensor Ising model in statistical physics is a generalization of the Ising model to $r$-order interactions. \cite{MukherjeeEtAl2022} study the one-parameter discrete exponential family for modeling dependent binary data where the interaction structure is given. In \cite{LiuEtAl2023} the tensor structure itself is to be inferred. These models are fundamentally different from our approach where we rely on properties of the quadratic exponential family which models up to second-order interactions. Another important difference is that we adopt the multi-linear formulation as it is inherently linked to the observable structure of multi-way data as opposed to describing the model coefficients with an $r$-order tensor structure.  
The $p$-dimensional Ising model is a discrete probability distribution on the set of $p$-dimensional binary vectors $\mat{x}\in\{0, 1\}^p$ with probability mass function (pmf) given by
\begin{displaymath}
    P_{\mat{\gamma}}(\mat{x}) = p_0(\mat{\gamma})\exp(\t{\vech(\mat{x}\t{\mat{x}})}\mat{\gamma}).
\end{displaymath}
The scaling factor $p_0(\mat{\gamma})\in\mathbb{R}_{+}$ ensures that $P_{\mat{\gamma}}$ is a pmf. It is equal to the probability of the zero event: $P(X = \mat{0}) = P_{\mat{\gamma}}(\mat{0}) = p_0(\mat{\gamma})$. More commonly known as the \emph{partition function}, the reciprocal of $p_0$, is given by
\begin{equation}\label{eq:ising-partition-function}
    p_0(\mat{\gamma})^{-1} = \sum_{\mat{x}\in\{0, 1\}^p}\exp(\t{\vech(\mat{x}\t{\mat{x}})}\mat{\gamma}).
\end{equation}
Abusing notation, let $\mat{\gamma}_{j l}$ denote the element of $\mat{\gamma}$ corresponding to $\mat{x}_j\mat{x}_l$ in $\vech(\mat{x}\t{\mat{x}})$.\footnote{Specifically, the element $\mat{\gamma}_{j l}$ of $\mat{\gamma}$ is a short hand for $\mat{\gamma}_{\iota(j, l)}$ with $\iota(j, l) = (\min(j, l) - 1)(2 p - \min(j, l)) / 2 + \max(j, l)$ mapping the matrix row index $j$ and column index $l$ to the corresponding half vectorization indices $\iota(j, l)$.} The ``diagonal'' parameter $\mat{\gamma}_{j j}$ expresses the conditional log odds of $X_j = 1\mid X_{-j} = \mat{0}$, where the negative subscript in $X_{-j}$ describes the $p - 1$ dimensional vector $X$ with the $j$th element removed. The off diagonal entries $\mat{\gamma}_{j l}$, $j\neq l$, are equal to the conditional log odds of simultaneous occurrence $X_j = 1, X_l = 1 \mid X_{-j, -l} = \mat{0}$. More precisely, the conditional probabilities $\pi_j(\mat{\gamma}) = P_{\mat{\gamma}}(X_j = 1\mid X_{-j} = \mat{0})$ and $\pi_{j, l}(\mat{\gamma}) = P_{\mat{\gamma}}(X_j = 1, X_l = 1\mid X_{-j, -l} = \mat{0})$ are related to the natural parameters via
\begin{equation}\label{eq:ising-two-way-log-odds}
    \mat{\gamma}_{j j} = \log\frac{\pi_j(\mat{\gamma})}{1 - \pi_j(\mat{\gamma})}, \qquad
    \mat{\gamma}_{j l} = \log\frac{1 - \pi_j(\mat{\gamma})\pi_l(\mat{\gamma})}{\pi_j(\mat{\gamma})\pi_l(\mat{\gamma})}\frac{\pi_{j l}(\mat{\gamma})}{1 - \pi_{j l}(\mat{\gamma})}.
\end{equation}

Conditional Ising models, incorporating the information of covariates $Y$ into the model, were considered by \cite{ChengEtAl2014,BuraEtAl2022}. The direct way is to parameterize $\mat{\gamma} = \mat{\gamma}_y$ by the covariate $Y = y$ to model a conditional distribution $P_{\mat{\gamma}_y}(\mat{x}\mid Y = y)$.

We extend the conditional pmf by allowing the binary variables to be tensor-valued; that is, we set $\mat{x} = \vec{\ten{X}}$, with dimension $p = \prod_{k = 1}^{r}p_k$ for $\ten{X}\in\{ 0, 1 \}^{p_1\times\cdots\times p_r}$. The tensor structure of $\ten{X}$ is accommodated by assuming Kronecker product constraints to the parameter vector $\mat{\gamma}_y$ in a similar fashion as in the multi-linear normal model. This means that we compare the pmf $P_{\mat{\gamma}_y}(\vec{\ten{X}} | Y = y)$ with the quadratic exponential family \eqref{eq:quad-density} with the natural parameters modeled by \eqref{eq:eta1} and \eqref{eq:eta2}. The diagonal of $(\vec{\ten{X}})\t{(\vec{\ten{X}})}$ is equal to $\vec{\ten{X}}$, which results in the GMLM being expressed as
\begin{align}
    P_{\mat{\gamma}_y}(\ten{X} \mid Y = y)
        &= p_0(\mat{\gamma}_y)\exp(\t{\vech((\vec{\ten{X}})\t{(\vec{\ten{X}})})}\mat{\gamma}_y) \label{eq:ising-cond-prob} \\
        &= p_0(\mat{\gamma}_y)\exp\Bigl(\Bigl\langle \ten{X}, \ten{F}_y\mlm_{k = 1}^{r}\mat{\beta}_k \Bigr\rangle + \Bigl\langle\ten{X}\mlm_{k = 1}^{r}\mat{\Omega}_k, \ten{X}\Bigr\rangle\Bigr) \nonumber
\end{align}
where we set $\overline{\ten{\eta}} = 0$. This imposes an additional constraint on the model, as the diagonal elements of $\mat{\Omega} = \bigkron_{k = r}^{1}\mat{\Omega}_k$ take the role of $\overline{\ten{\eta}}$, although not fully. Based on this modeling choice, the relation between the natural parameters $\mat{\gamma}_y$ of the conditional Ising model and the GMLM parameters $\mat{\beta}_k$ and $\mat{\Omega}_k$ is
\begin{equation}\label{eq:ising-natural-params}
    \mat{\gamma}_y
        = \t{\mat{D}_p}\vec(\mat{\Omega} + \diag(\mat{B}\vec{\ten{F}_y}))
        = \t{\mat{D}_p}\vec\Biggl(\bigkron_{k = r}^{1}\mat{\Omega}_k + \diag\biggl(\vec\Bigl(\ten{F}_y\mlm_{k = 1}^{r}\mat{\beta}_k\Bigr)\biggr)\Biggr).
\end{equation}
In contrast to the multi-linear normal GMLM, the matrices $\mat{\Omega}_k$ are only required to be symmetric. More specifically, we require $\mat{\Omega}_k$, for $k = 1, \ldots, r$, to be elements of an embedded submanifold of $\SymMat{p_k}$ (see \cref{sec:kron-manifolds,sec:matrix-manifolds}). The mode-wise reduction matrices $\mat{\beta}_k$ are elements of an embedded submanifold of $\mathbb{R}^{p_k\times q_k}$. Common choices are listed in \cref{sec:matrix-manifolds}.

To solve the optimization problem \eqref{eq:mle}, given a data set $(\ten{X}_i, y_i)$, $i = 1, \ldots, n$, we use the gradients in \cref{thm:grad} and a variation of gradient descent as described in \cref{sec:ising-param-estimation}, which is mostly technical in nature.



%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Statistical Properties}\label{sec:statprop}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
% \subsection{Asymptotics}\label{sec:asymtotics}
% %%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

Let $Z$ be a random variable %distributed according to a parameterized probability distribution 
with density $f_{\mat{\theta_0}}\in\{ f_{\mat{\theta}}: \mat{\theta}\in\Theta \}$, where $\Theta$ is a subset of a Euclidean space. We want to estimate the parameter that indexes the pdf,  ${\mat{\theta}}_0$, using $n$ i.i.d. (independent and identically distributed) copies of $Z$. We assume a known, real-valued and measurable function $m_{\mat{\theta}}$ exists with $z\mapsto m_{\mat{\theta}}(z)$ for every $\mat{\theta}\in\Theta$, and that  ${\mat{\theta}}_0$ is the unique maximizer of the map $\mat{\theta}\mapsto M(\mat{\theta}) = \E m_{\mat{\theta}}(Z)$. For the estimation, we maximize the empirical version
\begin{align}\label{eq:Mn}
    M_n(\mat{\theta}) &= \frac{1}{n}\sum_{i = 1}^n m_{\mat{\theta}}(Z_i).
\end{align}
An \emph{M-estimator} $\hat{\mat{\theta}}_n = \hat{\mat{\theta}}_n(Z_1, ..., Z_n)$ is a maximizer for the objective function $M_n$ over the parameter space $\Theta$ defined as
\begin{displaymath}
    \hat{\mat{\theta}}_n = \argmax_{\mat{\theta}\in\Theta} M_n(\mat{\theta}).
\end{displaymath}
If  the objective function $M_n$ is the log-likelihood,  $\widehat{\mat{\theta}}_n$ is the MLE. In general, the following asymptotic theory is valid for more than just the MLE. Moreover, it is not even necessary to have a \textit{perfect} maximizer, as long as the objective has finite supremum, it is sufficient to take an \emph{almost maximizer} $\hat{\mat{\theta}}_n$ as defined in the following.

\begin{definition}[weak and strong M-estimators]
    An estimator $\hat{\mat{\theta}}_n$ for the objective function $M_n$ in \eqref{eq:Mn} with $\sup_{\mat{\theta}\in\Theta}M_n(\mat{\theta}) < \infty$ such that
    \begin{displaymath}
        M_n(\hat{\mat{\theta}}_n) \geq \sup_{\mat{\theta}\in\Theta}M_n(\mat{\theta}) - o_P(n^{-1})
    \end{displaymath}
    is called a \emph{strong M-estimator} over $\Theta$. Replacing $o_P(n^{-1})$ by $o_P(1)$ gives a \emph{weak M-estimator}.
\end{definition}

\begin{theorem}[\hyperlink{proof:asymptotic-normality-gmlm}{Asymptotic Normality}]\label{thm:asymptotic-normality-gmlm}
    Assume $Z = (\ten{X}, Y)$ satisfies model \eqref{eq:quad-density} subject to \eqref{eq:eta1-manifold} and \eqref{eq:eta2-manifold} with true constrained parameter $\mat{\theta}_0 = (\overline{\eta}_0, \mat{B}_0, \vech{\mat{\Omega}_0})\in\Theta$, where $\Theta$ is defined in \cref{thm:param-manifold}. Under the regularity \crefrange{cond:differentiable-and-convex}{cond:finite-sup-on-compacta} in  \cref{app:proofs}, there exists a strong M-estimator sequence $\hat{\mat{\theta}}_n$ deriving from $l_n$ in \eqref{eq:log-likelihood} over $\Theta$. Furthermore, any strong M-estimator $\hat{\mat{\theta}}_n$ converges in probability to the true parameter $\mat{\theta}_0$, $\hat{\mat{\theta}}_n\xrightarrow{p}\mat{\theta}_0$, over $\Theta$. Moreover, every strong M-estimator $\hat{\mat{\theta}}_n$ is asymptotically normal,
    \begin{displaymath}
        % \sqrt{n}(\hat{\mat{\theta}}_n - \mat{\theta}_0) \xrightarrow{d} \mathcal{N}_{p(p + 2 q + 3) / 2}(0, \mat{\Sigma}_{\mat{\theta}_0})
        \sqrt{n}(\hat{\mat{\theta}}_n - \mat{\theta}_0) \xrightarrow{d} \mathcal{N}(0, \mat{\Sigma}_{\mat{\theta}_0})
    \end{displaymath}
    with asymptotic variance-covariance  $\mat{\Sigma}_{\mat{\theta}_0}$ given in \eqref{eq:asymptotic-covariance-gmlm} in \cref{app:proofs}.
\end{theorem}

To provide an intuition for the asymptotic variance-covariance structure $\mat{\Sigma}_{\mat{\theta}_0}$, we start from the classical, non-degenerate setting of an MLE $\hat{\mat{\xi}}_n$ in an unconstrained parameter space $\Xi$ containing the true parameter $\mat{\xi}_0$. In this case, $\mat{\Sigma}_{\mat{\xi}_0}$ is symmetric positive definite. Such matrices can be associated with a hyper-ellipsoid with axes associated with the eigenvectors of $\mat{\Sigma}_{\mat{\xi}_0}$. Given the manifold parameter space $\Theta\subseteq\Xi$ with true parameter $\mat{\theta}_0 = \mat{\xi}_0$, the asymptotic variance-covariance $\mat{\Sigma}_{\mat{\theta}_0}$ is a positive semi-definite matrix associated with a (degenerate) hyper-ellipsoid resulting from intersecting the hyper-ellipsoid of $\mat{\Sigma}_{\mat{\xi}_0}$ with the tangent space of $\Theta$ at $\mat{\theta}_0 = \mat{\xi}_0$ and distorting its shape with respect to the local curvature of $\Theta$ at $\mat{\theta}_0$.

\begin{remark}
    \cref{thm:asymptotic-normality-gmlm} is a special case of a more general asymptotic normality \cref{thm:M-estimator-asym-normal-on-manifolds} that also generalizes Theorem~5.23 in \cite{vanderVaart1998}, where $\Theta$ is an open subset of an Euclidean space, which is the simplest form of an embedded manifold. \Cref{thm:M-estimator-asym-normal-on-manifolds} is provided in \cref{app:proofs} due to its technical nature.
\end{remark}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Simulations}\label{sec:simulations}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
In this section, we report simulation results for the multi-linear normal and the multi-linear Ising model where different aspects of the GMLM model are compared against other methods. These are: \textit{Tensor Sliced Inverse Regression} (TSIR) \cite[]{DingCook2015},  an extension of Sliced Inverse Regression (SIR) \cite{Li1991} to tensor-valued predictors; the \textit{Multiway Generalized Canonical Correlation Analysis} (MGCCA) \cite[]{ChenEtAl2021,GirkaEtAl2024}, an extension of canonical correlation analysis (CCA) designed to handle multi-block data with tensor  structure; and the Tucker decomposition that is a higher-order form of principal component analysis (HOPCA) \cite[]{KoldaBader2009}, for both continuous and binary data. For the latter, the binary values are treated as continuous. As part of our baseline analysis, we also incorporate traditional Principal Component Analysis (PCA) on vectorized observations. In the case of the Ising model, we also compare with LPCA (Logistic PCA) and CLPCA (Convex Logistic PCA), both introduced in \cite{LandgrafLee2020}. All experiments are performed with sample sizes $n = 100, 200, 300, 500$ and $750$. Each experiment is repeated $100$ times.

To assess the accuracy of the estimation of  $\ten{R}(\ten{X})$ in \cref{thm:sdr}, we compare the estimate with the true vectorized reduction matrix $\mat{B} = \bigkron_{k = r}^{1}\mat{\beta}_k$, as it is compatible with any linear reduction method. We compute the \emph{subspace distance}, $d(\mat{B}, \hat{\mat{B}})$, between  $\mat{B}\in\mathbb{R}^{p\times q}$ and an estimate $\hat{\mat{B}}\in\mathbb{R}^{p\times \tilde{q}}$, which satisfies 
\begin{displaymath}
    d(\mat{B}, \hat{\mat{B}}) \propto \| \mat{B}\pinv{(\t{\mat{B}}\mat{B})}\t{\mat{B}} - \hat{\mat{B}}\pinv{(\t{\hat{\mat{B}}}\hat{\mat{B}})}\t{\hat{\mat{B}}} \|_F,
\end{displaymath}
where $\propto$ signifies proportional to. %the Frobenius norm of the difference between the projections onto the span of $\mat{B}$ and $\hat{\mat{B}}$. 
The proportionality constant\footnote{The proportionality constant depends on the dimension $p$ and the ranks of $\mat{B}$ and $\hat{\mat{B}}$. The explicit value of the proportionality constant is given by $(\min(\rank\mat{B} + \rank\hat{\mat{B}}, 2 p - (\rank\mat{B} + \rank\hat{\mat{B}})))^{-1/2}$.}  ensures  $d(\mat{B}, \hat{\mat{B}}) \in [0, 1]$. A distance of zero implies space overlap and a distance of one implies orthogonality of the subspaces.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Multi-Linear Normal}\label{sec:sim-tensor-normal}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
We generate a random sample $y_i$, $i=1,\ldots, n$,  from the standard normal distribution. We then draw independent samples $\ten{X}_i$  from the conditional multi-linear normal distribution of $\ten{X}\mid Y = y_i$ for $i = 1, ..., n$. The conditional distribution $\ten{X}\mid Y = y_i$ depends on the choice of the GMLM parameters $\overline{\ten{\eta}}$, $\mat{\beta}_1, ..., \mat{\beta}_r$, $\mat{\Omega}_1, ..., \mat{\Omega}_r$, and the function $\ten{F}_y$ of $y$. In all experiments we set $\overline{\ten{\eta}} = \mat{0}$. The other parameters and $\ten{F}_y$ are described per experiment. Given the true GMLM parameters and $\ten{F}_y$, we compute the conditional multi-linear normal mean $\ten{\mu}_y = \ten{F}_y\mlm_{k = 1}^{r}\mat{\Omega}_k^{-1}\mat{\beta}_k$ and covariances $\mat{\Sigma}_k = \mat{\Omega}_k^{-1}$ as in \eqref{eq:tnormal_cond_params}.

We consider the following settings:
\begin{itemize}
    \item[1a)] $\ten{X}$ is a three-way ($r = 3$) array of dimension  $2\times 3\times 5$, and $\ten{F}_y\equiv y$ is a $1\times 1\times 1$ tensor. The true $\mat{\beta}_k$'s are all equal to $\mat{e}_1\in\mathbb{R}^{p_k}$, the first unit vector, for $k \in \{1, 2, 3\}$. The matrices $\mat{\Omega}_k$ have an auto-regression like structure ($\mathrm{AR}(0.5)$), with entries $(\mat{\Omega}_k)_{i j} = 0.5^{|i - j|}$.
    \item[1b)] $\ten{X}$ is a three-way ($r = 3$) array of dimension  $2\times 3\times 5$, and relates to the response $y$ via a qubic polynomial. This is modeled via $\ten{F}_y$ of dimension $2\times 2\times 2$ by the twice iterated outer product of the vector $(1, y)$, with elements $(\ten{F}_y)_{i j k} = y^{i + j + k - 3}$. All $\mat{\beta}_k$'s are set to $(\mat{e}_1, \mat{e}_2)\in\mathbb{R}^{p_k\times 2}$ with $\mat{e}_i$ the $i$th unit vector and the $\mat{\Omega}_k$'s are $\mathrm{AR}(0.5)$.
    \item[1c)] Same as 1b), except that the GMLM parameters $\mat{\beta}_k$ are rank $1$ given by
    \begin{displaymath}
        \mat{\beta}_1 = \begin{pmatrix} 1 & -1 \\ -1 & 1 \end{pmatrix},\quad
        \mat{\beta}_2 = \begin{pmatrix} 1 & -1 \\ -1 & 1 \\ 1 & -1 \end{pmatrix},\quad
        \mat{\beta}_3 = \begin{pmatrix} 1 & -1 \\ -1 & 1 \\ 1 & -1 \\ -1 & 1 \\ 1 & -1 \end{pmatrix}.
    \end{displaymath}
    \item[1d)] Same as 1b), but  the true $\mat{\Omega}_k$ is tri-diagonal with elements $(\mat{\Omega}_k)_{i j} = \delta_{0, |i - j|} + 0.5\delta_{1, |i - j|}$, where $\delta_{i, j}$ is the Kronecker delta, for $k = 1, 2, 3$. 
    
    \item[1e)] For the misspecification model, we let $\ten{X}\mid Y$ be multivariate  but \emph{not} multi-linear normal. Let $\ten{X}$ be a $5\times 5$ random matrix with normal entries, $Y$ a univariate standard normal and $\mat{f}_y = (1, \sin(y), \cos(y), \sin(y)\cos(y))$. The true vectorized reduction matrix $\mat{B}$ is $25\times 4$ consisting of the first $4$ columns of the identity; i.e., $B_{i j} = \delta_{i j}$. The variance-covariance matrix $\mat{\Sigma}$ has elements $\Sigma_{i j} = 0.5^{|i - j|}$. Both, $\mat{B}$ and $\mat{\Omega} = \mat{\Sigma}^{-1}$ violate the Kronecker product assumptions \eqref{eq:eta1} and \eqref{eq:eta2} of the GMLM model. We set
    \begin{displaymath}
        \vec{\ten{X}}\mid (Y = y) = \mat{B}\mat{f}_y + \mathcal{N}_{25}(\mat{0}, \mat{\Sigma}),
    \end{displaymath}
    but we fit the model with the wrong  $\ten{F}_y$: We set $\ten{F}_y$ to be a $2\times 2$ matrix with $(\ten{F}_y)_{i j} = y^{i + j - 2}$, $i,j=1,2$. Experiment 1e) involves a deliberately misspecified model designed to evaluate the robustness of our approach. %The true model does \emph{not} have a Kronecker structure and the ``known'' function $\ten{F}_y$ of $y$ is misspecified as well.
\end{itemize}



\begin{figure}[H]
    \centering
    \includegraphics[scale = 1]{../plots/sim-normal-2x3.pdf}
    \caption{\label{fig:sim-normal}Multi-linear normal GMLM. The mean  subspace distance $d(\mat{B}, \hat{\mat{B}})$ over $100$ replications is plotted versus sample size on the $x$-axis. The simulation settings are described in \cref{sec:sim-tensor-normal}.}
\end{figure}

We plot the subspace distance between the true and estimated $\mat{B}$ over sample size in \cref{fig:sim-normal}. For 1a), TSIR and GMLM are equivalent and the best performers, %as expected given a 1D linear relation between the response $Y$ and $\ten{X}$. 
as expected. \cite{DingCook2015} had established that TSIR obtains the MLE estimate under a multi-linear (matrix) normal distributed setting. MGCCA, HOPCA and PCA do not even come close with MGCCA being slightly better than PCA which, unexpectedly, beats HOPCA.  
Continuing with 1b), where we introduced a cubic relation between $Y$ and $\ten{X}$, GMLM performs slightly better than TSIR. This is  mainly due to GMLM  estimating an $8$ dimensional subspace, which amplifies the small performance boost we gain by avoiding slicing.  The GMLM model in 1c) behaves as expected, clearly being the best. The other results are surprising. First, PCA, HOPCA and MGCCA are visually indistinguishable. This is explained by a high signal-to-noise ratio in this particular example. The most unexpected outcome is the failure of TSIR, especially because the conditional distribution $\ten{X}\mid Y$ is multi-linear normal, which, combined with $\cov(\vec\ten{X})$ exhibiting a Kronecker structure, should yield the MLE estimate. The low-rank assumption poses no challenges, as it corresponds to TSIR estimating a one-dimensional linear reduction. % that satisfies all the necessary criteria. 
A well-known issue with slicing methods, as applied in TSIR, is that conditional multi-modal distributions may lead to estimation problems due to varying distribution modes causing slice means to vanish. However, this issue does not arise in simulation 1c). An investigation into this behavior revealed the problem lies in the estimation of the mode covariance matrices $\mat{O}_k = \E[(\ten{X} - \E\ten{X})_{(k)}\t{(\ten{X} - \E\ten{X})_{(k)}}]$. The mode-wise reductions in TSIR are computed as $\hat{\mat{O}}_k^{-1}\hat{\mat{\Gamma}}_k$. Poor estimation of $\mat{O}_k$ causes the failure of TSIR. This is due to the high signal-to-noise ratio in this particular simulation. GMLM exhibits excellent performance in high signal-to-noise ratio settings but may be less robust in low signal-to-noise ratio settings where TSIR can perform better, at least in this specific example. Simulation 1d), incorporating information about the covariance structure behaves similarly to 1b), except that GMLM gains a statistically significant lead in estimation performance over TSIR. The last simulation, 1e), where the model is misspecified and GMLM is at a theoretical disadvantage, all three GMLM, TSIR, and MGCCA, are on par. GMLM demonstrates a modest advantage up to a sample size of 300, whereas MGCCA takes the lead after a sample size of 500. PCA and HOPCA are both outperformed.% \textcolor{red}{A wrong assumption about the relation to the response is still better than no relation at all.}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Ising Model}\label{sec:sim-ising}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
 We let $Y_i$ be i.i.d. uniform on $[-1,1]$, $i = 1, \ldots, n$,  and $\ten{X}$ be $2\times 3$ %dimensional binary matrix 
with conditional matrix (multi-linear) Ising distribution $\ten{X}\mid Y$ as in \cref{sec:ising_estimation}.  Unless otherwise specified, we set the GMLM parameters to be % $\mat{\beta}_1, \mat{\beta}_2$, $\mat{\Omega}_1, \mat{\Omega}_2$. We let
\begin{displaymath}
    \mat{\beta}_1 = \begin{pmatrix}
        1 & 0 \\ 0 & 1
    \end{pmatrix}, \quad \mat{\beta}_2 = \begin{pmatrix}
        1 & 0 \\ 0 & 1 \\ 0 & 0
    \end{pmatrix}, \quad \mat{\Omega}_1 = \begin{pmatrix}
        0 & -2 \\ -2 & 0
    \end{pmatrix}, \quad \mat{\Omega}_2 = \begin{pmatrix}
        1 & 0.5 & 0 \\
        0.5 & 1 & 0.5 \\
        0 & 0.5 & 1
    \end{pmatrix}
\end{displaymath}
and
\begin{displaymath}
    \ten{F}_y = \begin{pmatrix}
        \sin(\pi y) & -\cos(\pi y) \\
        \cos(\pi y) & \sin(\pi y)
    \end{pmatrix}.
\end{displaymath}

We consider the settings: 
\begin{itemize}
    \item[2a)] A purely linear relation between $\mat{X}$ and the response with  $\ten{F}_y\equiv y:1\times 1$ and $\t{\mat{\beta}_1} = (1, 0)$ and $\t{\mat{\beta}_2} = (1, 0, 0)$.
    \item[2b)] The ``baseline'' simulation with all parameters as described above.
    \item[2c)] Low rank regression with both $\mat{\beta}_1$ and $\mat{\beta}_2$ of rank $1$,
    \begin{displaymath}
        \mat{\beta}_1 = \begin{pmatrix}
            1 & 0 \\ 1 & 0
        \end{pmatrix}, \qquad \mat{\beta}_2 = \begin{pmatrix}
            0 & 0 \\ 1 & -1 \\ 0 & 0
        \end{pmatrix}.
    \end{displaymath}
    \item[2d)] %We conclude with a simulation relating to 
    The original design of the Ising model is a mathematical model of the behavior of Ferromagnetism \cite{Ising1925} in a thermodynamic setting, modeling the interaction effects of elementary magnets (spin up/down relating to $0$ and $1$). The model assumes all elementary magnets to be the same, which translates to all having the same coupling strength (two-way interactions) governed by a single parameter relating to the temperature of the system. Assuming the magnets are organized in a 2D grid represented by matrix-valued $\ten{X}$, their interactions are restricted to direct neighbors. This is modeled by setting the true $\mat{\Omega}_k$s as tri-diagonal matrices with zero diagonal entries and all non-zero entries identical. %Since this is a 1D matrix manifold, we can enforce the constraint. Setting the true interaction parameters to be
    We set
    \begin{displaymath}
        \mat{\Omega}_1 = \frac{1}{2}\begin{pmatrix}
            0 & 1 \\ 1 & 0
        \end{pmatrix}, \qquad \mat{\Omega}_2 = \begin{pmatrix}
            0 & 1 & 0 \\
            1 & 0 & 1 \\
            0 & 1 & 0
        \end{pmatrix}
    \end{displaymath}
    where $1 / 2$ corresponds to an arbitrary temperature. The mean effect depending on $\ten{F}_y$ can be interpreted as an external magnetic field.
\end{itemize}

\begin{figure}[ht!]
    \centering
    \includegraphics[scale = 1]{../plots/sim-ising.pdf}
    \caption{\label{fig:sim-ising}Simulation results for the Ising GMLM. Sample sizes are placed on the $x$-axis and the mean of subspace distance $d(\mat{B}, \hat{\mat{B}})$ over $100$ replications on the $y$-axis.}
\end{figure}

The subspace distances between the estimated and true parameter spaces for sample size $n=100, 200, 300, 500, 750$ are plotted in \cref{fig:sim-ising}. 
The comparative results are similar across all simulation settings. PCA and HOPCA, both treating the response $\ten{X}$ as continuous, perform poorly. Not much better are LPCA and CLPCA. Similar to PCA and HOPCA, they do not account for the relation with the response, but they are specifically created for binary predictors.  MGCCA, which accounts for $y$ in reducing the predictors, outperforms all the PCA variants. TSIR comes next, even though it treats the predictors $\ten{X}$ as continuous. The Ising GMLM model is the best performer in all simulations.

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Data Analysis}\label{sec:data-analysis}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
We apply GMLM and competing methods to the EEG data set. The data. which are freely available\footnote{\url{http://kdd.ics.uci.edu/databases/eeg/eeg.data.html} (accessed April 29, 2025)}, were partially described in \cref{sec:introduction}. Each subject was exposed to either a single stimulus (S1) or to two stimuli (S1 and S2) which were pictures of objects chosen from the 1980 Snodgrass and Vanderwart picture set (see data documentation). When two stimuli were shown, they were presented in either a matched condition where S1 was similar to S2 or in a non-matched condition where S1 differed from S2. These different stimulus conditions introduce a third dimension, making the data a 3-tensor with dimension of the third axis $p_3 = 3$. It is common in the analysis of this data set to only consider the S1 stimulus by dropping all but the S1 setting as described in \cref{sec:introduction} in analogy to \cite{LiKimAltman2010,PfeifferForzaniBura2012,DingCook2015,PfeifferKaplaBura2021}. Here we consider both cases, denoted ether the 2D EEG data set (matrix-valued) for the S1 stimulus only in contrast to the 3D EED data set (tensor-valued) incorporating all stimuli scenarios.

SDR methods for regression with matrix-valued predictors were developed by \cite{LiKimAltman2010} (folded SIR), \cite{PfeifferForzaniBura2012} (L(ongitudinal)SIR), \cite{DingCook2014} (dimension folding PCA and PFC), \cite{PfeifferKaplaBura2021} (K-PIR (ls), K-PIR (mle)), and \cite{DingCook2015} (T(ensor)SIR). LSIR is a generalization of the simple SIR nonparametric slicing algorithm \cite{Li1991} to matrix-valued regressors. The K-PIR (ls and mle) algorithm is based on regressing the matrix-valued predictors on the response using a bilinear regression model and can be considered as a continuous parametric modeling extension of LSIR. TSIR generalizes matrix-valued SDR to tensor-valued SDR. In this data set, $p = p_1 p_2 = 16384$ is much larger than $n=122$. To deal with this issue, the predictor data were pre-screened via (2D)$^2$PCA \cite[]{ZhangZhou2005} which reduced the dimension of the predictor matrix to $(p_1, p_2) = (3, 4)$, $(15, 15)$ and $(20, 30)$ \cite[]{PfeifferKaplaBura2021, LiKimAltman2010, DingCook2014, DingCook2015}. \cite{PfeifferKaplaBura2021} also carried out simultaneous dimension reduction and variable selection using the fast POI-C algorithm of \cite{JungEtAl2019} without data pre-processing\footnote{Due to high computational burden, only 10-fold cross-validation was performed for fast POI-C}.

For the 3D EEG version of the data only TSIR and GMLM are directly applicable since the other methods are specifically designed for the matrix case or require significantly bigger sample sizes. Nevertheless, both K-PIR and LSIR are easily generalized from matrix-valued to tensor-valued predictors. In addition, it is possible to apply LSIR to the full data set without pre-screening from an algorithmic point of few, ignoring any theory or assumptions. Only K-PIR can not be applied to the full data set.

In contrast, TSIR and our GMLM model can be applied directly to the raw data without pre-screening or variable selection. In general, the sample size does not need to be large for maximum likelihood estimation in the multilinear normal model. In particular for matrix normal models, \cite{DrtonEtAl2020} proved that very small sample sizes, as little as $3$,\footnote{The required minimum sample size depends on non-trivial number theoretic relations between the mode dimensions, while the vectorized dimensionality has no specific role.} are sufficient to obtain unique MLEs for Kronecker covariance structures.
Among the tensor-focused methods in \cref{sec:simulations}, we herein consider only TSIR in view of its superior performance.

We report results for the best performing methods from \cite{PfeifferKaplaBura2021}, namely K-PIR (ls) and LSIR from \cite{PfeifferForzaniBura2012} for $(p_1, p_2) = (3, 4)$, $(15, 15)$, and $(20, 30)$. In addition, we report results for GMLM and two versions of TSIR. TSIR in its original form \cite[]{DingCook2015} as well as TSIR (reg), where we introduce a similar regularization as the one used in the tensor-normal GMLM algorithm. Regularizing TSIR was deemed appropriate in view of the surprising 2D EEG results for TSIR (see \cref{sec:GMLMvsTSIR} for a comparison of TSIR versus GMLM leading to TSIR (reg)). We report results from applying  GMLM as well as GMLM (FFT) (Fast Fourier Transform) %without any parameter constraints while GMLM (FFT)
which assumes that the time-axis of the EEG data contains important underlying frequencies. For this reason, the time-axis reduction is modeled as a time-series mixture containing only the $10$ frequencies of highest magnitude as lower-magnitude ones often correspond to noise or less important variations in the signal. %Ignoring them can improve the signal-to-noise ratio. %This modeling assumption is fully supported by our GMLM as described in  \cref{sec:manifolds}.
The discriminatory ability of the reduced predictors of the various approaches was assessed by AUC or the area under the receiver operator characteristic (ROC) curve \cite[p. 67]{Pepe2003}. We use leave-one-out cross-validation to obtain unbiased AUC estimates. The average AUC values and their standard deviation for the 2D and 3D EEG data are tabulated in \cref{tab:eeg}.

% Across all pre-screening dimensions, K-PIR (ls) has an AUC of $78\%$. LSIR performs better at the price of some instability; it peaked at $85\%$ at $(3, 4)$, then dropped down to $81\%$ at $(15, 15)$, and then increased to $83\%$.  In contrast, our GMLM method peaked at $(3, 4)$ with $85\%$ and stayed stable at $84\%$, also at the full data ($(p_1, p_2) = (256, 64)$). In contrast,  fast POI-C that carries out simultaneous feature extraction and feature selection is outperformed by all other methods with an AUC of $63\%$. Folded SIR \cite[]{LiKimAltman2010} is excluded as it exhibits consistently lower AUC values, ranging from 0.61 to 0.70. \daniel{Tensor SIR to the table, still needs description in text.}

Across all pre-screening dimensions, K-PIR (ls) has an AUC of $78\%$ in the 2D EEG with worse results in the 3D case with an AUC of $74\%$--$75\%$. This is in stark contrast to all other methods which gained predictive performance from 2D to 3D. LSIR showed excellent performance with high pre-screening, peaking at $85\%$ and $88\%$ in the 2D and 3D setting, respectively. Without pre-screening, LSIR fails, which is expected as it does not support the full data setting where the time-axis dimension exceeds the sample size, regardless of its algorithmic applicability. 

In the 2D scenario, TSIR shows equally good performance with high pre-screening at $(4, 3)$ while slowly decreasing in performance down to $80\%$ at $(20, 30)$ and then dropping down to $69\%$ AUC for the full data set. Comparing with the TSIR performance in the 3D case is somewhat surprising as it is quite stable over all pre-screening dimensions as well as on the full data. This inconsistency between the 2D and 3D analysis of TSIR in conjunction with an expectation of TSIR behaving similarly to GMLM in this case steered us to take a closer look in \cref{sec:GMLMvsTSIR}. We deduced that the main reason for the noted unexpected performance of TSIR is caused by the required inversion of ill-conditioned variance-covariance estimates. To address this limitation, we amended it to TSIR (reg) by introducing regularization of the variance-covariance matrix. %if it is ill-conditioned yielded 
TSIR (reg), as can be seen in \cref{tab:eeg}, is very stable with better performance in the 2D case compared to TSIR and a slight decrease in the 3D setting. Our GMLM has identical performance to TSIR (reg) in the 2D setting, as we expected. In the 3D case, GMLM is slightly better than  TSIR (reg) while maintaining stability in performance across the different settings. Finaly, GMLM (FFT), which makes only sense to be applied to the full data, reaches AUC of $85\%$ in 2D and $87\%$ in 3D.

% \begin{table}[!hpt]
%     \caption{\label{tab:eeg}Mean AUC values and their standard deviation (in parentheses) based on leave-one-out cross-validation for the 2D EEG imaging data (77 alcoholic and 45 control subjects)}
%     \begin{tabular}{l l c c c}
%     \hline
%     Method  & \multicolumn{4}{c}{$(p_1, p_2)$} \\
%             & \multicolumn{1}{c}{$(3, 4)$}
%             & \multicolumn{1}{c}{$(15, 15)$}
%             & \multicolumn{1}{c}{$(20, 30)$}
%             & \multicolumn{1}{c}{$(256, 64)$} \\
%     \hline
%     % K-PIR (ls) & 0.78 (0.04) & 0.78 (0.04) & 0.78 (0.04) & \\
%     % LSIR       & 0.85 (0.04) & 0.81 (0.04) & 0.83 (0.04) & \\
%     % TSIR       & 0.85 (0.04) & 0.83 (0.04) & 0.80 (0.04) & 0.69 (0.05) \\
%     % TSIR (reg) & 0.85 (0.04) & 0.84 (0.04) & 0.84 (0.04) & 0.84 (0.04) \\
%     % FastPOI-C  & & & & 0.63 (0.22)\makebox[0pt]{\ \ ${}^{*}$} \\
%     % GMLM       & 0.85 (0.04) & 0.84 (0.04) & 0.84 (0.04) & 0.84 (0.04) \\
%     K-PIR (ls) & 78\% (4\%) & 78\% (4\%) & 78\% (4\%) & \\
%     LSIR       & 85\% (4\%) & 81\% (4\%) & 83\% (4\%) & 39\% (5\%) \\
%     TSIR       & 85\% (4\%) & 83\% (4\%) & 80\% (4\%) & 69\% (5\%) \\
%     TSIR (reg) & 85\% (4\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
%     FastPOI-C  & & & & 63\% (22\%)\makebox[0pt]{\ \ ${}^{*}$} \\
%     GMLM       & 85\% (4\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
%     GMLM (FFT) & & & & 85\% (4\%) \\
%     \hline
%     \multicolumn{4}{l}{${}^{*}$\footnotesize based on 10-fold cross-validation.}
%     \end{tabular}
% \end{table}
%
% \begin{table}[!hpt]
%     \centering
%     \caption{\label{tab:eeg3d}Mean AUC values and their standard deviation (in parentheses) based on leave-one-out cross-validation for the 3D EEG imaging data (77 alcoholic and 45 control subjects)}
%     \begin{tabular}{l l c c c}
%         \hline
%         Method  & \multicolumn{4}{c}{$(p_1, p_2, p_3)$} \\
%                 & \multicolumn{1}{c}{$(3, 4, 3)$}
%                 & \multicolumn{1}{c}{$(15, 15, 3)$}
%                 & \multicolumn{1}{c}{$(20, 30, 3)$}
%                 & \multicolumn{1}{c}{$(256, 64, 3)$} \\
%         \hline
%         K-PIR (ls) & 74\% (5\%) & 75\% (4\%) & 75\% (4\%) \\
%         LSIR       & 88\% (3\%) & 81\% (4\%) & 71\% (5\%) & 52\% (6\%) \\
%         TSIR       & 87\% (3\%) & 89\% (3\%) & 88\% (3\%) & 86\% (3\%) \\
%         TSIR (reg) & 87\% (3\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
%         GMLM       & 88\% (3\%) & 87\% (3\%) & 87\% (3\%) & 87\% (3\%) \\
%         GMLM (FFT) & & & & 87\% (3\%) \\
%         \hline
%     \end{tabular}
% \end{table}

\begin{table}[!hpt]
    \caption{\label{tab:eeg}Mean AUC values and their standard deviation (in parentheses) based on leave-one-out cross-validation for the EEG imaging data}
    \begin{tabular}{l c c c c}
    \hline
            & \multicolumn{4}{c}{2D EEG: $(p_1, p_2)$} \\
    Method  & \multicolumn{1}{c}{$(3, 4)$}
            & \multicolumn{1}{c}{$(15, 15)$}
            & \multicolumn{1}{c}{$(20, 30)$}
            & \multicolumn{1}{c}{$(256, 64)$} \\
    \hline
    K-PIR (ls) & 78\% (4\%) & 78\% (4\%) & 78\% (4\%) & \\
    LSIR       & 85\% (4\%) & 81\% (4\%) & 83\% (4\%) & 39\% (5\%) \\
    TSIR       & 85\% (4\%) & 83\% (4\%) & 80\% (4\%) & 69\% (5\%) \\
    TSIR (reg) & 85\% (4\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
    % FastPOI-C  & & & & 63\% (22\%)\makebox[0pt]{\ \ ${}^{*}$} \\
    GMLM       & 85\% (4\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
    GMLM (FFT) & & & & 85\% (4\%) \\
    \hline
            & \multicolumn{4}{c}{3D EEG: $(p_1, p_2, p_3)$} \\
    Method  & \multicolumn{1}{c}{$(3, 4, 3)$}
            & \multicolumn{1}{c}{$(15, 15, 3)$}
            & \multicolumn{1}{c}{$(20, 30, 3)$}
            & \multicolumn{1}{c}{$(256, 64, 3)$} \\
    \hline
    K-PIR (ls) & 74\% (5\%) & 75\% (4\%) & 75\% (4\%) \\
    LSIR       & 88\% (3\%) & 81\% (4\%) & 71\% (5\%) & 52\% (6\%) \\
    TSIR       & 87\% (3\%) & 89\% (3\%) & 88\% (3\%) & 86\% (3\%) \\
    TSIR (reg) & 87\% (3\%) & 84\% (4\%) & 84\% (4\%) & 84\% (4\%) \\
    GMLM       & 88\% (3\%) & 87\% (3\%) & 87\% (3\%) & 87\% (3\%) \\
    GMLM (FFT) & & & & 87\% (3\%) \\
    \hline
    % \multicolumn{4}{l}{${}^{*}$\footnotesize based on 10-fold cross-validation.}
    \end{tabular}
\end{table}


As a real data application on regressions with binary tensor-valued predictors, we analyze chess data and perform a proof of concept data analysis where a chess board is interpreted as a collection of binary $8\times 8$ matrices in \cref{sec:chess}. 

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Discussion}\label{sec:discussion}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
In this paper, we propose a generalized multi-linear model formulation for the inverse conditional distribution of a tensor-valued predictor given a response and derive a multi-linear sufficient reduction for the corresponding forward regression/classification problem. We also propose estimators for the sufficient reduction and show they are consistent and asymptotically normal. We demonstrate, through a numerical example in \cref{sec:GMLMvsVecEfficiency}, the modeling benefits of leveraging the tensor structure of the data.

Obtaining the asymptotic results required leveraging manifolds as a basis for resolving the issue of unidentifiable parameters. This in turn led to an even more flexible modeling framework, which allows building complex and potentially problem-specific parameter spaces that incorporate additional domain-specific knowledge into the model.
We allude to this feature of our approach in \cref{sec:matrix-manifolds}, where we also tabulate different matrix manifolds that can be used as building blocks $\manifold{B}_k$ and $\manifold{O}_k$ of the parameter space in \cref{tab:matrix-manifolds}. For example, our formulation can easily accommodate longitudinal data tabulated in matrix format, where the rows are covariates and the columns are consecutive time points with discrete AR($k$) dependence structure. 

Our multi-linear Ising model can be thought of as the extension of the Ising model-based approach of  \cite{ChengEtAl2014}, where a $q$-dimensional binary vector is regressed on a $p$-dimensional continuous vector. Yet, our model leverages the inherent structural information of the tensor-valued covariates by assuming separable first and second moments. By doing so, it bypasses requiring sparsity assumptions or penalization, despite the tensor high-dimensional nature of the data. Moreover, it can accommodate a mixture of continuous and binary tensor-valued predictors, which is the subject of future work.  

A special case of the Ising model is the one-parameter Ising model without an external field. Given our theory, it is possible to represent this special case as well by constructing a one-dimensional parameter manifold. More precisely, if $\mat{A}$ is a fixed symmetric $p\times p$ matrix with zero main diagonal elements,  the parameter space for the one-dimensional Ising model has the form $\Theta = \{\mat{0}\}\times \{ \theta\mat{A} + 0.5\mat{I}_p : \theta\in\mathbb{R} \}$. As a result (a) the model is a vector-valued multi-variate binary model for predictors $\mat{X}\in\{0, 1\}^p$, and (b) the predictors $\mat{X}$ are independent of the response. In the unsupervised setting, the usual question is to determine the single parameter $\theta$. This is the model considered by \cite{XuMukherjee2023}, whose finding that the asymptotic distribution of the MLE depends on the true value of the parameter and is not necessarily normal, at first glance, may seem contradictory to our asymptotic result that the MLE $\hat{\theta}_n$ is asymptotically normal. \cite{XuMukherjee2023} answer a different question in that they determine the limiting experiment distribution \cite[][Ch~9]{vanderVaart1998},  where they let the dimension $p$ of the binary vector-valued predictors  go to infinity ($p\to\infty$). We, on the other hand, let the sample size; i.e., the number of observations go to infinity ($n\to\infty$).

An additional powerful extension of our model involves considering a sum of separable Kronecker predictors. This is motivated by the equivalence of a Kronecker product to a rank $1$ tensor. By allowing a sum of a few separable Kronecker predictors, we remove the implicit rank $1$ constraint. However, if this extension is to be applied to the SDR setting, as in this paper, it is crucial to ensure that the sum of Kronecker products form a parameter manifold.



%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
%% Multiple Appendixes:                     %%
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\begin{appendix}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Examples}\label{app:examples}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

\begin{example}[Vectorization]\label{ex:vectorization}
    Given a matrix
    \begin{displaymath}
        \mat{A} = \begin{pmatrix}
            1 & 4 & 7 \\
            2 & 5 & 8 \\
            3 & 6 & 9
        \end{pmatrix}
    \end{displaymath}
    its vectorization is $\vec{\mat{A}} = \t{(1, 2, 3, 4, 5, 6, 7, 8, 9)}$ and its half vectorization $\vech{\mat{A}} = \t{(1, 2, 3, 5, 6, 9)}$. Let $\ten{A}$ be a tensor of dimension $3\times 3\times 3$ given by
    \begin{displaymath}
        \ten{A}_{:,:,1} = \begin{pmatrix}
            1 & 4 & 7 \\
            2 & 5 & 8 \\
            3 & 6 & 9
        \end{pmatrix},
        \qquad
        \ten{A}_{:,:,2} = \begin{pmatrix}
            10 & 13 & 16 \\
            11 & 14 & 17 \\
            12 & 15 & 18
        \end{pmatrix},
        \qquad
        \ten{A}_{:,:,3} = \begin{pmatrix}
            19 & 22 & 25 \\
            20 & 23 & 26 \\
            21 & 24 & 27
        \end{pmatrix}.
    \end{displaymath}
    Then the vectorization of $\ten{A}$ is given by
    \begin{displaymath}
        \vec{\ten{A}} = \t{(1, 2, 3, 4, ..., 26, 27)}\in\mathbb{R}^{27}.
    \end{displaymath}
\end{example}

\begin{example}[Matricization]
    Let $\ten{A}$ be the $3\times 4\times 2$ tensor given by
    \begin{displaymath}
        \ten{A}_{:,:,1} = \begin{pmatrix}
            1 & 4 & 7 & 10 \\
            2 & 5 & 8 & 11 \\
            3 & 6 & 9 & 12
        \end{pmatrix},
        \ten{A}_{:,:,2} = \begin{pmatrix}
            13 & 16 & 19 & 22 \\
            14 & 17 & 20 & 23 \\
            15 & 18 & 21 & 24
        \end{pmatrix}.
    \end{displaymath}
    Its matricizations are then
    \begin{gather*}
        \ten{A}_{(1)} = \begin{pmatrix}
            1 &  4 &  7 & 10 & 13 & 16 & 19 & 22 \\
            2 &  5 &  8 & 11 & 14 & 17 & 20 & 23 \\
            3 &  6 &  9 & 12 & 15 & 18 & 21 & 24
        \end{pmatrix},
        \qquad
        \ten{A}_{(2)} = \begin{pmatrix}
             1 &  2 &  3 & 13 & 14 & 15 \\
             4 &  5 &  6 & 16 & 17 & 18 \\
             7 &  8 &  9 & 19 & 20 & 21 \\
            10 & 11 & 12 & 22 & 23 & 24
        \end{pmatrix}, \\
        \ten{A}_{(3)} = \begin{pmatrix}
             1 &  2 &  3 &  4 &  5 &  6 &  7 &  8 &  9 &  10 &  11 &  12 \\
            13 & 14 & 15 & 16 & 17 & 18 & 19 & 20 & 21 &  22 &  23 &  24
        \end{pmatrix}.
    \end{gather*}
\end{example}


% %%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
% \section{Multi Linear Algebra}\label{app:multi-linear-algebra}
% %%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
%
% \begin{displaymath}
%     (\ten{A}\circ\ten{B})\mlm_{k = 1}^{r + s} \mat{C}_k
%     =
%     \Bigl(\ten{A}\mlm_{k = 1}^r \mat{C}_k\Bigr)\circ\Bigl(\ten{B}\mlm_{l = 1}^s \mat{C}_{l + r}\Bigr)
% \end{displaymath}
% Using $\K(\ten{A}\circ\ten{B}) = \ten{A}\otimes\ten{B}$ gives
% \begin{displaymath}
%     \K\Bigl((\ten{A}\circ\ten{B})\mlm_{k = 1}^{r + s} \mat{C}_k\Bigr)
%     =
%     \Bigl(\ten{A}\mlm_{k = 1}^r \mat{C}_k\Bigr)\otimes\Bigl(\ten{B}\mlm_{l = 1}^s \mat{C}_{l + r}\Bigr)
% \end{displaymath}
% A generalization of the well-known identity $\vec(\mat{A}\mat{B}\mat{C}) = (\t{\mat{C}}\otimes\mat{A})\vec{\mat{B}}$ is given by
% \begin{displaymath}
%     \Bigl(\ten{A}\mlm_{k = 1}^r \mat{B}_k \Bigr)_{(\mat{i}, \mat{j})}
%     =
%     \Bigl( \bigotimes_{k = \#\mat{i}}^{1}\mat{B}_{\mat{i}_k} \Bigr)
%         \ten{A}_{(\mat{i}, \mat{j})}
%             \Bigl( \bigotimes_{l = \#\mat{j}}^{1}\t{\mat{B}_{\mat{j}_l}} \Bigr)
% \end{displaymath}
% with the special case
% \begin{displaymath}
%     \vec\Bigl(\ten{A}\mlm_{k = 1}^r \mat{B}_k\Bigr)
%     =
%     \Bigl(\bigotimes_{k = r}^{1}\mat{B}_k\Bigr)\vec{\ten{A}}
% \end{displaymath}

% Furthermore, we have
% \begin{displaymath}
%     (\ten{A}\otimes\ten{B})\mlm_{k = 1}^{r}\t{(\vec\mat{C}_k)}
%     =
%     \Bigl\langle \ten{A}\mlm_{k = 1}^{r} \mat{C}_k, \ten{B} \Bigr\rangle
%     =
%     \Bigl\langle \ten{A}, \ten{B}\mlm_{k = 1}^{r} \t{\mat{C}_k} \Bigr\rangle
%     =
%     \t{(\vec{\ten{B}})}\Bigl(\bigotimes_{k = r}^{1}\mat{C}_k\Bigr)\vec{\ten{A}}
% \end{displaymath}
% as well as for any tensor $\ten{A}$ of even order $2 r$ and matching square matrices $\mat{B}_k$ holds
% \begin{displaymath}
%     \K(\ten{A})\mlm_{k = 1}^{r}\t{(\vec\mat{B}_k)}
%     =
%     \t{(\vec{\ten{A}})}\vec\Bigl(\bigotimes_{k = r}^{1}\t{\mat{B}_k}\Bigr)
% \end{displaymath}

% \begin{lemma}\label{thm:kron-perm}
%     Given $r \geq 2$ matrices $\mat{A}_k$ of dimension $p_j\times q_j$ for $k = 1, \ldots, r$, then there exists a unique permutation matrix $\mat{S}_{\mat{p}, \mat{q}}$ such that
%     \begin{equation}\label{eq:kron-to-outer-perm}
%         \vec\bigkron_{k = r}^{1}\mat{A}_k = \mat{S}_{\mat{p}, \mat{q}}\vec\bigouter_{k = 1}^{r}\mat{A}_k.
%     \end{equation}
%     The permutation $\mat{S}_{\mat{p}, \mat{q}}$ with indices $\mat{p} = (p_1, \ldots, p_r)$ and $\mat{q} = (q_1, \ldots, q_r)$ is defined recursively as
%     \begin{equation}\label{eq:S_pq}
%         \mat{S}_{\mat{p}, \mat{q}} = \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)} \bigl(\mat{I}_{p_r q_r}\otimes\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\bigr)
%     \end{equation}
%     with initial value
%     \begin{displaymath}
%         \mat{S}_{(p_1, p_2), (q_1, q_2)} = \mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1}
%     \end{displaymath}
%     where $\mat{K}_{p, q}$ is the \emph{commutation matrix} \cite[][Ch.~11]{AbadirMagnus2005}, that is the permutation such that $\vec{\t{\mat{A}}} = \mat{K}_{p, q}\vec{\mat{A}}$ for every $p\times q$ dimensional matrix $\mat{A}$.
% \end{lemma}
% \begin{proof}
%     Lemma~7 in \cite{MagnusNeudecker1986} states that
%     \begin{align}
%         \vec(\mat{A}_2\otimes\mat{A}_1)
%             &= (\mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1})(\vec{\mat{A}_2}\otimes\vec{\mat{A}_1}) \label{eq:MagnusNeudecker1986-vec-kron-identity} \\
%             &= (\mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1})\vec(\mat{A}_1\circ \mat{A}_2). \nonumber
%     \end{align}
%     This proves the statement for $r = 2$. The general statement for $r > 2$ follows via induction. Assuming \eqref{eq:kron-to-outer-perm} holds for $r - 1$, the induction step is then;
%     \begin{multline*}
%         \vec{\bigkron_{k = r}^{1}}\mat{A}_k
%         = \vec\Bigl(\mat{A}_r\otimes\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr) \\
%         \overset{\eqref{eq:MagnusNeudecker1986-vec-kron-identity}}{=} \Bigl( \mat{I}_{q_r}\otimes\mat{K}_{\prod_{k = 1}^{r - 1}q_k, p_r}\otimes\mat{I}_{\prod_{k = 1}^{r - 1}p_k} \Bigr)\vec\Bigl((\vec\mat{A}_r)\otimes\vec\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr) \\
%         = \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)}\vec\Bigl[\Bigl(\vec\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
%         \overset{\eqref{eq:kron-to-outer-perm}}{=} \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)}\vec\Bigl[\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\Bigl(\vec\bigouter_{k = 1}^{r - 1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
%         \overset{(a)}{=} \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)} \bigl(\mat{I}_{p_r q_r}\otimes\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\bigr)\vec\Bigl[\Bigl(\vec\bigouter_{k = 1}^{r - 1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
%         = \mat{S}_{\mat{p}, \mat{q}}\vec\bigouter_{k = 1}^{r}\mat{A}_k.
%     \end{multline*}
%     Equality $(a)$ uses the relation $\vec(\mat{C}\mat{a}\t{\mat{b}}) = (\mat{I}_{\dim(\mat{b})}\otimes\mat{C})\vec(\mat{a}\t{\mat{b}})$ for a matrix $\mat{C}$ and vectors $\mat{a}, \mat{b}$.
% \end{proof}

% \begin{remark}
%     The permutation matrix $\mat{K}_{p, q}$ represents a perfect outer $p$-shuffle of $p q$ elements.
% \end{remark}



%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Proofs}\label{app:proofs}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

\begin{proof}[\hypertarget{proof:sdr}{Proof of \cref{thm:sdr}}]
    A direct implication of Theorem~1 from \cite{BuraDuarteForzani2016} is that, under the exponential family \eqref{eq:quad-density} with natural statistic $\mat{t}(\ten{X})$,
    \begin{displaymath}
        \t{\mat{\alpha}}(\mat{t}(\ten{X}) - \E\mat{t}(\ten{X}))
    \end{displaymath}
    is a sufficient reduction, where $\mat{\alpha}\in\mathbb{R}^{p (p + 3) / 2\times q}$ with $\Span(\mat{\alpha}) = \Span(\{\mat{\eta}_y - \E_{Y}\mat{\eta}_Y : y\in\mathcal{S}_Y\})$ where $\mathcal{S}_Y$ denotes the sample space of $Y$. Since $\E_Y\ten{F}_Y = 0$ and $\E_Y\mat{\eta}_{1 Y} = \E[\vec\overline{\ten{\eta}} - \mat{B}\vec\ten{F}_Y] = \vec\overline{\ten{\eta}}$ is follows that
    \begin{displaymath}
        \mat{\eta}_y - \E_{Y}\mat{\eta}_Y = \begin{pmatrix}
            \mat{\eta}_{1 y} - \E_{Y}\mat{\eta}_{1 Y} \\
            \mat{\eta}_{2} - \E_{Y}\mat{\eta}_{2}
        \end{pmatrix} = \begin{pmatrix}
            \mat{B}\vec\ten{F}_y \\
            \mat{0}
        \end{pmatrix}.
    \end{displaymath}
    because $\mat{\eta}_{2}$ does not depend on $y$. The set $\{ \vec{\ten{F}_y} : y\in\mathcal{S}_Y \}$ is a subset of $\mathbb{R}^q$. Therefore,
    \begin{displaymath}
        \Span\left(\{\mat{\eta}_y - \E_{Y}\mat{\eta}_Y : y\in\mathcal{S}_Y\}\right) = \Span\left(\left\{\begin{pmatrix}
            \mat{B}\vec\ten{F}_Y \\ \mat{0}
        \end{pmatrix} : y\in\mathcal{S}_Y \right\}\right)
        \subseteq
        \Span\begin{pmatrix}
            \mat{B} \\ \mat{0}
        \end{pmatrix},
    \end{displaymath}
    which obtains with $\mat{t}(\ten{X}) = (\vec\ten{X}, \vech(\t{(\vec\ten{X})}(\vec\ten{X})))$ that 
    \begin{displaymath}
        \t{\begin{pmatrix}
            \mat{B} \\ \mat{0}
        \end{pmatrix}}(\mat{t}(\ten{X}) - \E\mat{t}(\ten{X}))
        =
        \t{\mat{B}}\vec(\ten{X} - \E\ten{X})
        = \vec\Bigl(\ten{F}_y\mlm_{k = 1}^{r}\mat{\beta}_k\Bigr)
    \end{displaymath}
    is also a sufficient reduction, though not necessarily minimal, using $\mat{B} = \bigkron_{k = 1}^{r}\mat{\beta}_k$. When the exponential family is full rank, which in our setting amounts to all $\mat{\beta}_j$'s being full rank matrices for $j = 1, \ldots, r$, then Theorem~1 from \cite{BuraDuarteForzani2016} also obtains the minimality of the reduction.
\end{proof}

\begin{proof}[\hypertarget{proof:kron-manifolds}{Proof of \cref{thm:kron-manifolds}}]
    We start by considering the first case and assume that $\manifold{B}$ is spherical with radius $1$ w.l.o.g. We equip $\manifold{K} = \{ \mat{A}\otimes \mat{B} : \mat{A}\in\manifold{A}, \mat{B}\in\manifold{B} \}\subset\mathbb{R}^{p_1 p_2\times q_1 q_2}$ with the subspace topology.\footnote{Given a topological space $(\manifold{M}, \mathcal{T})$ and a subset $\mathcal{S}\subseteq\manifold{M}$ the \emph{subspace topology} on $\manifold{S}$ induced by $\manifold{M}$ consists of all open sets in $\manifold{M}$ intersected with $\manifold{S}$, that is $\{ O\cap\manifold{S} : O\in\mathcal{T} \}$ \cite[]{Lee2012,Lee2018,Kaltenbaeck2021}.} Define the hemispheres $H_i^{+} = \{ \mat{B}\in\manifold{B} : (\vec{\mat{B}})_i > 0 \}$ and $H_i^{-} = \{ \mat{B}\in\manifold{B} : (\vec{\mat{B}})_i < 0 \}$ for $i = 1, ..., p_2 q_2$. The hemispheres are an open cover of $\manifold{B}$ with respect to the subspace topology. Define for every $H_i^{\pm}$, where $\pm$ is a placeholder for ether $+$ or $-$, the function
    \begin{displaymath}
        f_{H_i^{\pm}} : \manifold{A}\times H_i^{\pm}\to\mathbb{R}^{p_1 p_2\times q_1 q_2}
            : (\mat{A}, \mat{B})\mapsto \mat{A}\otimes \mat{B}
    \end{displaymath}
    which is smooth. With the spherical property of $\manifold{B}$ the relation $\|\mat{A}\otimes \mat{B}\|_F = \|\mat{A}\|_F$ for all $\mat{A}\otimes \mat{B}\in\manifold{K}$ ensures that the function $f_{H_i^{\pm}}$, constrained to its image, is bijective with inverse function (identifying $\mathbb{R}^{p\times q}$ with $\mathbb{R}^{p q}$) given by
    \begin{displaymath}
        f_{H_i^{\pm}}^{-1} = \begin{cases}
            f_{H_i^{\pm}}(\manifold{A}\times H_i^{\pm})\to\manifold{A}\times H_i^{\pm} \\
            \mat{C}\mapsto \left(\pm\frac{\|\mat{C}\|_F}{\|\mat{R}(\mat{C})\mat{e}_i\|_2}\mat{R}(\mat{C})\mat{e}_i, \pm\frac{1}{\|\mat{C}\|_F\|\mat{R}(\mat{C})\mat{e}_i\|_2}\mat{R}(\mat{C})\t{\mat{R}(\mat{C})}\mat{e}_i\right)
        \end{cases}
    \end{displaymath}
    where $\pm$ is $+$ for a ``positive'' hemisphere $H_i^{+}$ and $-$ otherwise, $\mat{e}_i\in\mathbb{R}^{p_2 q_2}$ is the $i$th unit vector and $\mat{R}(\mat{C})$ is a ``reshaping'' permutation\footnote{Relating to $\K$ the operation $\mat{R}$ is basically its inverse as $\K(\mat{A}\circ\mat{B}) = \mat{A}\otimes\mat{B}$ with a mismatch in the shapes only.} which acts on Kronecker products as $\mat{R}(\mat{A}\otimes \mat{B}) = (\vec{\mat{A}})\t{(\vec{\mat{B}})}$. This makes $f_{H_i^{\pm}}^{-1}$ a combination of smooth functions ($\mat{0}$ is excluded from $\manifold{A}, \manifold{B}$ guarding against division by zero) and as such it is also smooth. This ensures that $f_{H_i^{\pm}} : \manifold{A}\times {H_i^{\pm}}\to f_{H_i^{\pm}}(\manifold{A}\times {H_i^{\pm}})$ is a diffeomorphism.

    Next, we construct an atlas\footnote{A collection of charts $\{ \varphi_i : i\in I \}$ with index set $I$ of a manifold $\manifold{A}$ is called an \emph{atlas} if the pre-images of the charts $\varphi_i$ cover the entire manifold $\manifold{A}$.} for $\manifold{K}$ which is equipped with the subspace topology. Let $(\varphi_j, U_j)_{j\in J}$ be a atlas of $\manifold{A}\times\manifold{B}$. Such an atlas exists and admits a unique smooth structure as both $\manifold{A}, \manifold{B}$ are embedded manifolds from which we take the product manifold. The images of the coordinate domains $f_H(U_j)$ are open in $\manifold{K}$, since $f_H$ is a diffeomorphism, with the corresponding coordinate maps
    \begin{displaymath}
        \phi_{H_i^{\pm},j} : f_{H_i^{\pm}}(U_j)\to \varphi_j(U_j)
            : \mat{C}\mapsto \varphi_j(f_{H_i^{\pm}}^{-1}(\mat{C})).
    \end{displaymath}
    By construction the set $\{ \phi_{H_i^{\pm},j} : i = 1, ..., p_2 q_2, \pm\in\{+, -\}, j\in J \}$ is an atlas if the charts are compatible. This means we need to check if the transition maps are diffeomorphisms. Let $(\phi_{H, j}, V_j), (\phi_{\widetilde{H}, k}, V_k)$ be two arbitrary charts from our atlas, then the transition map $\phi_{\widetilde{H}, k}\circ\phi_{H,j}^{-1}:\phi_{H,j}^{-1}(V_j\cap V_k)\to\phi_{\widetilde{H},k}^{-1}(V_j\cap V_k)$ has the form
    \begin{displaymath}
        \phi_{\widetilde{H}, k}\circ\phi_{H,j}^{-1}
            = \varphi_k\circ f_{\widetilde{H}}^{-1}\circ f_{H}\circ\varphi_{j}^{-1}
            = \varphi_k\circ (\pm\mathrm{id})\circ\varphi_{j}^{-1}
    \end{displaymath}
    where $\pm$ depends on $H, \widetilde{H}$ being of the same ``sign'' and $\mathrm{id}$ is the identity. We conclude that the charts are compatible, which makes the constructed set of charts an atlas. With that we have shown the topological manifold $\manifold{K}$ with the subspace topology admit a smooth atlas that makes it an embedded smooth manifold with dimension equal to the dimension of the product topology $\manifold{A}\times\manifold{B}$; that is, $d = \dim\manifold{A} + \dim\manifold{B}$.

    It remains to show that the cone condition also admits a smooth manifold. $\manifold{K} = \{ \mat{A}\otimes \mat{B} : \mat{A}\in\manifold{A}, \mat{B}\in\widetilde{\manifold{B}} \}$, where $\widetilde{\manifold{B}} = \{ \mat{B}\in\manifold{B} : \|\mat{B}\|_F = 1 \}$, holds if both $\manifold{A}, \manifold{B}$ are cones. Since $g:\manifold{B}\to\mathbb{R}:\mat{B}\mapsto \|\mat{B}\|_F$ is continuous on $\manifold{B}$ with full rank $1$ everywhere,  $\widetilde{\manifold{B}} = g^{-1}(1)$ is a $\dim{\manifold{B}} - 1$ dimensional embedded submanifold of $\manifold{B}$. An application of the spherical case proves the cone case.
\end{proof}

\begin{proof}[\hypertarget{proof:param-manifold}{Proof of \cref{thm:param-manifold}}]
    Induction using \cref{thm:kron-manifolds} ensures that $\manifold{K}_{\mat{B}}$ and $\manifold{K}_{\mat{\Omega}}$ are embedded submanifolds.

    Since $\vech : \mathbb{R}^{p^2}\to\mathbb{R}^{p(p+1)/2}$ is linear and $\manifold{K}_{\mat{\Omega}}$ is embedded we get by the linearity of $\vech$ that $\vech(\manifold{K}_{\mat{\Omega}}) = \{\vech\mat{\Omega}:\mat{\Omega}\in\manifold{K}_{\mat{\Omega}}\}$ is embedded in $\mathbb{R}^{p(p+1)/2}$.

    Finally, the finite product manifold of embedded submanifolds is embedded in the finite product space of their ambient spaces, that is $\Theta = \mathbb{R}^p \times \manifold{K}_{\mat{B}}\times\vech(\manifold{K}_{\mat{\Omega}}) \subset \mathbb{R}^p\times\mathbb{R}^{p\times q}\times\mathbb{R}^{p(p + 1)/2}$ is embedded.
\end{proof}

\begin{proof}[\hypertarget{proof:grad}{Proof of \cref{thm:grad}}]
    We first note that for any exponential family with density \eqref{eq:quad-density} the term $b(\mat{\eta}_{y})$ differentiated with respect to the natural parameter $\mat{\eta}_{y}$ is the expectation of the statistic $\mat{t}(\ten{X})$ given $Y = y$. In our case, we get $\nabla_{\mat{\eta}_{y}}b = (\nabla_{\mat{\eta}_{1{y}}}b, \nabla_{\mat{\eta}_2}b)$ with components
    \begin{displaymath}
        \nabla_{\mat{\eta}_{1{y}}}b
            = \E[\vec(\ten{X})\mid Y = y]
            = \vec\E[\ten{X}\mid Y = y]
            % = \vec\ten{g}_1(\mat{\eta}_{y})
    \end{displaymath}
    and
    \begin{multline*}
        \nabla_{\mat{\eta}_{2}}b
           % &= \E[\mat{t}_2(\ten{X})\mid Y = y_i]
            = \E[\vech((\vec\ten{X})\t{(\vec\ten{X})})\mid Y = y] \\
            = \E[\pinv{\mat{D}_p}\vec(\ten{X}\circ\ten{X})\mid Y = y]
            % = \pinv{\mat{D}_p}\vec\ten{g}_2(\mat{\eta}_{y}).
            = \pinv{\mat{D}_p}\vec\E[\ten{X}\circ\ten{X}\mid Y = y].
    \end{multline*}
    The gradients are related to their derivatives by transposition, $\nabla_{\mat{\eta}_{1{y_i}}}b = \t{\D b(\mat{\eta}_{1{y}})}$ and $\nabla_{\mat{\eta}_2}b = \t{\D b(\mat{\eta}_2)}$.
    Next, we provide the differentials of the natural parameter components from \eqref{eq:eta1} and \eqref{eq:eta2} in a quite direct form, without any further ``simplifications,'' because the down-stream computations will not benefit from re-expressing
    \begin{align*}
        \d\mat{\eta}_{1{y}}(\overline{\ten{\eta}})
            &= \d\vec{\overline{\ten{\eta}}}, \\
        \d\mat{\eta}_{1{y}}(\mat{\beta}_j)
            &= \vec\Bigl( \ten{F}_{y}\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\beta}_k\times_j\d\mat{\beta}_j \Bigr), \\
        \d\mat{\eta}_2(\mat{\Omega}_j)
             = c\t{\mat{D}_p}\vec\d\bigotimes_{j = r}^1\mat{\Omega}_j
            &= c\t{\mat{D}_p}\vec\Bigl(\,\bigkron_{k = r}^{j + 1}\mat{\Omega}_k\otimes\d\mat{\Omega}_j\otimes\bigkron_{l = j - 1}^{1}\mat{\Omega}_l \Bigr).
    \end{align*}
    All other combinations, namely $\d\mat{\eta}_{1{y}}(\mat{\Omega}_j)$, $\d\mat{\eta}_2(\overline{\ten{\eta}})$ and $\d\mat{\eta}_2(\mat{\beta}_j)$, are zero.

    Continuing with the partial differentials of $l_n$ from \eqref{eq:log-likelihood}
    \begin{multline*}
        \d l_n(\overline{\ten{\eta}})
            = \sum_{i = 1}^{n} (\langle \d\overline{\ten{\eta}}, \ten{X}_i \rangle - \D b(\mat{\eta}_{1{y_i}})\d\mat{\eta}_{1{y_i}}(\overline{\ten{\eta}}))
            = \sum_{i = 1}^{n} \t{(\vec{\ten{X}_i} - \vec\E[\ten{X} \mid Y = y_i])}\d\vec{\overline{\ten{\eta}}} \\
            = \t{(\d\vec{\overline{\ten{\eta}}})}\vec\sum_{i = 1}^{n} (\ten{X}_i - \E[\ten{X} \mid Y = y_i]).
    \end{multline*}
    For every $j = 1, ..., r$ we get the differentials
    \begin{align*}
        \d l_n(\mat{\beta}_j)
             &= \sum_{i = 1}^{n} \biggl(\Bigl\langle \ten{F}_{y_i}\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\beta}_k\times_j\d\mat{\beta}_j, \ten{X}_i \Bigr\rangle - \D b(\mat{\eta}_{1{y_i}})\d\mat{\eta}_{1{y_i}}(\mat{\beta}_j)\biggr) \\
             &= \sum_{i = 1}^{n} \Bigl\langle \ten{F}_{y_i}\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\beta}_k\times_j\d\mat{\beta}_j, \ten{X}_i - \E[\ten{X} \mid Y = y_i] \Bigr\rangle \\
             &= \sum_{i = 1}^{n} \tr\biggl( \d\mat{\beta}_j\Bigl(\ten{F}_{y_i}\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\beta}_k\Bigr)_{(j)} \t{(\ten{X}_i - \E[\ten{X} \mid Y = y_i])_{(j)}} \biggr) \\
             &= \t{(\d\vec{\mat{\beta}_j})}\vec\sum_{i = 1}^{n} (\ten{X}_i - \E[\ten{X} \mid Y = y_i])_{(j)} \t{\Bigl(\ten{F}_{y_i}\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\beta}_k\Bigr)_{(j)}}
    \end{align*}
    as well as
    \begin{align*}
        \d l_n(\mat{\Omega}_j)
            &= \sum_{i = 1}^{n} \biggl( c\Bigl\langle \ten{X}_i\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\Omega}_k\times_j\d\mat{\Omega}_j, \ten{X}_i \Bigr\rangle - \D b(\mat{\eta}_2)\d\mat{\eta}_2(\mat{\Omega}_j) \biggr) \\
            &= c\sum_{i = 1}^{n} \biggl( \Bigl\langle \ten{X}_i\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\Omega}_k\times_j\d\mat{\Omega}_j, \ten{X}_i \Bigr\rangle \\
                &\qquad\qquad - \t{(\pinv{\mat{D}_p}\vec\E[\ten{X}\circ\ten{X}\mid Y = y_i])}\t{\mat{D}_p}\vec\Bigl(\,\bigkron_{k = r}^{j + 1}\mat{\Omega}_k\otimes\d\mat{\Omega}_j\otimes\bigkron_{l = j - 1}^{1}\mat{\Omega}_l \Bigr) \biggr) \\
            &\overset{(a)}{=} c\sum_{i = 1}^{n} \biggl( \Bigl\langle \ten{X}_i\mlm_{\substack{k = 1\\k\neq j}}^{r}\mat{\Omega}_k\times_j\d\mat{\Omega}_j, \ten{X}_i \Bigr\rangle \\
                &\qquad\qquad - \t{(\vec\E[\ten{X}\circ\ten{X}\mid Y = y_i])}\vec\Bigl(\,\bigkron_{k = r}^{j + 1}\mat{\Omega}_k\otimes\d\mat{\Omega}_j\otimes\bigkron_{l = j - 1}^{1}\mat{\Omega}_l \Bigr) \biggr) \\
            &= c\sum_{i = 1}^{n} \biggl( \t{\vec(\ten{X}_i\circ\ten{X}_i - \E[\ten{X}\circ\ten{X}\mid Y = y_i])}\vec\Bigl(\,\bigkron_{k = r}^{j + 1}\mat{\Omega}_k\otimes\d\mat{\Omega}_j\otimes\bigkron_{l = j - 1}^{1}\mat{\Omega}_l \Bigr) \biggr) \\
            &= c\sum_{i = 1}^{n} \bigl(\ten{X}_i\otimes\ten{X}_i - \E[\ten{X}\otimes\ten{X}\mid Y = y_i]\bigr)\mlm_{\substack{k = 1\\k\neq j}}^{r}\t{(\vec{\mat{\Omega}_k})}\times_j\t{(\d\vec{\mat{\Omega}_j})} \\
            &= c\t{(\d\vec{\mat{\Omega}_j})}\sum_{i = 1}^{n} \Bigl(\bigl(\ten{X}_i\otimes\ten{X}_i - \E[\ten{X}\otimes\ten{X}\mid Y = y_i]\bigr)\mlm_{\substack{k = 1\\k\neq j}}^{r}\t{(\vec{\mat{\Omega}_k})}\Bigr)_{(j)} \\
            &= c\t{(\d\vec{\mat{\Omega}_j})}\vec\sum_{i = 1}^{n} \bigl(\ten{X}_i\otimes\ten{X}_i - \E[\ten{X}\otimes\ten{X}\mid Y = y_i]\bigr)\mlm_{\substack{k = 1\\k\neq j}}^{r}\t{(\vec{\mat{\Omega}_k})}.
    \end{align*}
    Equality $(a)$ holds because $\mat{N}_p = \mat{D}_p\pinv{\mat{D}_p}$ is the symmetrizer matrix \cite[][Ch.~11]{AbadirMagnus2005}. The symmetrizer matrix $\mat{N}_p$ satisfies  $\mat{N}_p\vec{\mat{A}} = \vec{\mat{A}}$ if $\mat{A} = \t{\mat{A}}$, and
    \begin{displaymath}
        \vec\E[\ten{X}\circ\ten{X}\mid Y = y] = \vec\E[(\vec{\ten{X}})\t{(\vec{\ten{X}})}\mid Y = y]
    \end{displaymath}
    is the vectorization of a symmetric matrix.

    Finally, using the identity $\d \ten{A}(\ten{B}) = \t{(\d\vec{\ten{B}})}\nabla_{\ten{B}}\ten{A}$ identifies the partial gradients.
\end{proof}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Asymptotic Normality for M-estimators on Manifolds}
To proof asymptotic results, we need some technical lemmas as well as the definition of the tangent space of a manifold provided herein. We conclude this subsection with \cref{thm:M-estimator-asym-normal-on-manifolds}, a more general form of \cref{thm:asymptotic-normality-gmlm}.

The following is a reformulation of Lemma~2.3 from \cite{BuraEtAl2018} which assumes Condition~2.2 to hold. %The existence of a mapping in Condition~2.2 is not needed for Lemma~2.3. It suffices that the restricted parameter space $\Theta$ is a subset of the unrestricted parameter space $\Xi$, which is trivially satisfied in our setting. Under this, \cref{thm:exists-strong-M-estimator-on-subsets} follows directly from Lemma~2.3 in \cite{BuraEtAl2018}.
Condition~2.2 assumes the existence of a mapping from an open set into the unrestricted parameter space $\Xi$ such that the image of the mapping is the restricted parameter space $\Theta$. The existence of this mapping is not needed, instead the weaker assumption $\Theta\subseteq\Xi$ is sufficient with an almost identical proof.

\begin{lemma}[Existence of strong M-estimators on Subsets]\label{thm:exists-strong-M-estimator-on-subsets}
    Assume there exists a (weak/strong) M-estimator $\hat{\mat{\xi}}_n$ for $M_n$ over $\Xi$. Then, there exists a strong M-estimator $\hat{\mat{\theta}}_n$ for $M_n$ over any non-empty $\Theta\subseteq\Xi$.
\end{lemma}
\begin{proof}%[\hypertarget{proof:exists-strong-M-estimator-on-subsets}{Proof of \cref{thm:exists-strong-M-estimator-on-subsets}}]
    Let $\hat{\mat{\xi}}_n$ be a (weak/strong) M-estimator for the unconstrained problem. This gives by definition, in any case, that
    \begin{displaymath}
        \sup_{\mat{\xi}\in\Xi} M_n(\mat{\xi}) \leq M_n(\hat{\mat{\xi}}_n) + o_P(1).
    \end{displaymath}
    Since $\emptyset\neq\Theta\subseteq\Xi$, $\sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) \leq \sup_{\mat{\xi}\in\Xi} M_n(\mat{\xi})$ and with $M_n(\mat{\xi}) < \infty$ for any $\mat{\xi}\in\Xi$
    \begin{displaymath}
        P\Bigl( \sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) < \infty \Bigr)
        \geq
        P\Bigl( \sup_{\mat{\xi}\in\Xi} M_n(\mat{\xi}) < \infty \Bigr)
        \xrightarrow{n\to\infty}
        1.
    \end{displaymath}
    If $\sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) < \infty$, then, for any $0 < \epsilon_n$ there exists $\hat{\mat{\theta}}_n\in\Theta$ such that $\sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) - \epsilon_n \leq M_n(\hat{\mat{\theta}}_n)$. Therefore, we can choose $\epsilon_n\in o(n^{-1})$, which yields
    \begin{displaymath}
        P\Bigl( M_n(\hat{\mat{\theta}}_n) \geq \sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) - o(n^{-1}) \Bigr)
        \geq
        P\Bigl( \sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) < \infty \Bigr)
        \xrightarrow{n\to\infty}
        1.
    \end{displaymath}
    The last statement is equivalent to
    \begin{displaymath}
        M_n(\hat{\mat{\theta}}_n) \geq \sup_{\mat{\theta}\in\Theta} M_n(\mat{\theta}) - o_P(n^{-1}),
    \end{displaymath}
    which is the definition of $\hat{\mat{\theta}}_n$ being a strong M-estimator over $\Theta$.
\end{proof}



\begin{lemma}[Existence and Consistency of M-estimators on Subsets]\label{thm:M-estimator-consistency-on-subsets}
    Let $\Xi$ be a convex open subset of a Euclidean space and $\Theta\subseteq\Xi$ non-empty. Assume $\mat{\xi}\mapsto m_{\mat{\xi}}(z)$ is a strictly concave function on $\Xi$ for almost all $z$ and $z\mapsto m_{\mat{\xi}}(z)$ is measurable for all $\mat{\xi}\in\Xi$. Let $M(\mat{\xi}) = \E m_{\mat{\xi}}(Z)$ be a well defined function with a unique maximizer $\mat{\theta}_0\in\Theta\subseteq\Xi$; that is, $M(\mat{\theta}_0) > M(\mat{\xi})$ for all $\mat{\xi}\neq\mat{\theta}_0$. Also, assume 
    \begin{displaymath}
        \E\sup_{\mat{\xi}\in K}|m_{\mat{\xi}}(Z)| < \infty,
    \end{displaymath}
    for every non-empty compact $K\subset\Xi$.     Then, there exists a strong M-estimator $\hat{\mat{\theta}}_n$ of $M_n(\mat{\theta}) = \frac{1}{n}\sum_{i = 1}^{n} m_{\mat{\theta}}(Z_i)$ over the subset $\Theta$. Moreover, any strong M-estimator $\hat{\mat{\theta}}_n$ of $M_n$ over $\Theta$ converges in probability to $\mat{\theta}_0$, that is $\hat{\mat{\theta}}_n\xrightarrow{p}\mat{\theta}_0$.
\end{lemma}
\begin{proof}%[Proof of \cref{thm:M-estimator-consistency-on-subsets}]
    The proof follows the proof of Proposition~2.4 in \cite{BuraEtAl2018} with the same assumptions, except that we only require $\Theta$ to be a subset of $\Xi$. This is accounted for by replacing Lemma~2.3 in \cite{BuraEtAl2018} with \cref{thm:exists-strong-M-estimator-on-subsets} to obtain the existence of a strong M-estimator on $\Theta$.
\end{proof}


We also need the concept of a \emph{tangent space} to formulate asymptotic normality in a way that is independent of a particular coordinate representation. Intuitively, the tangent space at a point $\mat{x}\in\manifold{A}$ of the manifold $\manifold{A}$ is the hyperspace of all velocity vectors $\t{\nabla\gamma(0)}$ of any curve $\gamma:(-1, 1)\to\manifold{A}$ passing through $\mat{x} = \gamma(0)$, see \cref{fig:torus}. Locally, at $\mat{x} = \gamma(0)$ with a chart $\varphi$ we can write $\gamma(t) = \varphi^{-1}(\varphi(\gamma(t)))$ that gives $\Span\t{\nabla\gamma(0)} \subseteq \Span\t{\nabla\varphi^{-1}(\varphi(\mat{x}))}$. Taking the union over all smooth curves through $\mat{x}$ gives equality. The following definition leverages the simplified setup of smooth manifolds in Euclidean space.

\begin{figure}[!hpt]
    \centering
    \includegraphics[width = 0.5\textwidth]{../images/TorustangentSpace.pdf}
    \caption{\label{fig:torus}Visualization of the tangent space $T_{\mat{x}}\manifold{A}$ at $\mat{x}$ of the torus $\manifold{A}$. The torus $\manifold{A}$ is a 2-dimensional embedded manifold in $\mathbb{R}^3$. The tangent space $T_{\mat{x}}\manifold{A}\subset\mathbb{R}^3$ is a 2-dimensional hyperplane visualized with its origin $\mat{0}$ shifted to $\mat{x}$. Two curves $\gamma_1, \gamma_2$ with $\mat{x} = \gamma_1(0) = \gamma_2(0)$ are drawn on the torus. The curve velocity vectors $\t{\nabla\gamma_1(0)}$ and $\t{\nabla\gamma_2(0)}$ are drawn as tangent vectors with root $\mat{x}$.}
\end{figure}

\begin{definition}[Tangent Space]\label{def:tangent-space}
    Let $\manifold{A}\subseteq\mathbb{R}^p$ be an embedded smooth manifold and $\mat{x}\in\manifold{A}$. The \emph{tangent space} at $\mat{x}$ of $\manifold{A}$ is defined as
    \begin{displaymath}
        T_{\mat{x}}\manifold{A} := \Span\t{\nabla\varphi^{-1}(\varphi(\mat{x}))}
    \end{displaymath}
    for any chart $\varphi$ with $\mat{x}$ in the pre-image of $\varphi$.
\end{definition}

\Cref{def:tangent-space} is consistent since it can be shown that two different charts at the same point have identical span.

We are now ready to provide and prove \cref{thm:M-estimator-asym-normal-on-manifolds}, a more general version of \cref{thm:asymptotic-normality-gmlm}.

\begin{theorem}[\hyperlink{proof:M-estimator-asym-normal-on-manifolds}{Asymptotic Normality for M-estimators on Manifolds}]\label{thm:M-estimator-asym-normal-on-manifolds}
    Let $\Theta\subseteq\mathbb{R}^p$ be a smooth embedded manifold. For each $\mat{\theta}$ in a neighborhood in $\mathbb{R}^p$ of the true parameter $\mat{\theta}_0\in\Theta$ let $z\mapsto m_{\mat{\theta}}(z)$ be measurable and $\mat{\theta}\mapsto m_{\mat{\theta}}(z)$ be differentiable at $\mat{\theta}_0$ for almost all $z$. Assume also that there exists a measurable function $u$ such that $\E[u(Z)^2] < \infty$, and for almost all $z$ as well as all $\mat{\theta}_1, \mat{\theta}_2$ in a neighborhood of $\mat{\theta}_0$ such that
    \begin{displaymath}
        | m_{\mat{\theta}_1}\!(z) - m_{\mat{\theta}_2}\!(z) | \leq u(z) \| \mat{\theta}_1 - \mat{\theta}_2 \|_2.
    \end{displaymath}
    Moreover, assume that $\mat{\theta}\mapsto\E[m_{\mat{\theta}}(Z)]$ admits a second-order Taylor expansion at $\mat{\theta}_0$ in a neighborhood of $\mat{\theta}_0$ in $\mathbb{R}^p$ with a non-singular Hessian $\mat{H}_{\mat{\theta}_0} = \nabla^2_{\mat{\theta}}\E[m_{\mat{\theta}}(Z)]|_{\mat{\theta} = \mat{\theta}_0}\in\mathbb{R}^{p\times p}$.

    If $\hat{\mat{\theta}}_n$ is a strong M-estimator of $\mat{\theta}_0$ in $\Theta$, then $\hat{\mat{\theta}}_n$ is asymptotically normal
    \begin{displaymath}
        \sqrt{n}(\hat{\mat{\theta}}_n - \mat{\theta}_0) \xrightarrow{d} \mathcal{N}_p\left(\mat{0}, \mat{\Pi}_{\mat{\theta}_0} \E\left[\nabla_{\mat{\theta}} m_{\mat{\theta}_0}(Z)\t{(\nabla_{\mat{\theta}} m_{\mat{\theta}_0}(Z))}\right] \mat{\Pi}_{\mat{\theta}_0}\right)
    \end{displaymath}
    where $\mat{\Pi}_{\mat{\theta}_0} = \mat{P}_{\mat{\theta}_0}\pinv{(\t{\mat{P}_{\mat{\theta}_0}}\mat{H}_{\mat{\theta}_0}\mat{P}_{\mat{\theta}_0})}\t{\mat{P}_{\mat{\theta}_0}}$ and $\mat{P}_{\mat{\theta}_0}$ is any matrix whose span is the tangent space $T_{\mat{\theta}_0}\Theta$ of $\Theta$ at $\mat{\theta}_0$.
\end{theorem}
\begin{proof}[\hypertarget{proof:M-estimator-asym-normal-on-manifolds}{Proof of \cref{thm:M-estimator-asym-normal-on-manifolds}}]
    Let $\varphi:U\to\varphi(U)$ be a coordinate chart\footnote{By \cref{def:manifold}, the chart $\varphi : U\to\varphi(U)$ is bi-continuous, infinitely often continuously differentiable, and has a continuously differentiable inverse $\varphi^{-1} : \varphi(U) \to U$. Furthermore, the domain $U$ is open with respect to the trace topology on $\Theta$, which means that there exists an open set $O\subseteq\mathbb{R}^p$ such that $U = \Theta\cap O$.} with $\mat{\theta}_0\in U\subseteq\Theta$. As $\varphi$ is continuous, $\varphi(\hat{\mat{\theta}}_n)\xrightarrow{p}\varphi(\mat{\theta}_0)$ by the continuous mapping theorem on metric spaces \cite[][Theorem~18.11]{vanderVaart1998}, which implies $P(\varphi(\hat{\mat{\theta}}_n)\in\varphi(U))\xrightarrow{n\to\infty}1$.

    The next step is to apply Theorem~5.23 in \cite{vanderVaart1998} to $\hat{\mat{s}}_n = \varphi(\hat{\mat{\theta}}_n)$. Therefore, assume that $\hat{\mat{s}}_n\in\varphi(U)$. Denote with $\mat{s} = \varphi(\mat{\theta})\in\varphi(U)\subseteq\mathbb{R}^d$ the coordinates of the parameter $\mat{\theta}\in U\subseteq\Theta$ of the $d = \dim(\Theta)$ dimensional manifold $\Theta\subseteq\mathbb{R}^p$. Since  $\varphi : U\to\varphi(U)$ is bijective, we can express $m_{\mat{\theta}}$ in terms of $\mat{s} = \varphi(\mat{\theta})$  as $m_{\mat{\theta}} = m_{\varphi^{-1}(\mat{s})}$ for every $\mat{\theta}\in U$. Furthermore, we let
    \begin{displaymath}
        M(\mat{\theta}) = \E[m_{\mat{\theta}}(Z)] \qquad\text{and}\qquad M_{\varphi}(\mat{s}) = \E[m_{\varphi^{-1}(\mat{s})}(Z)] = M(\varphi^{-1}(\mat{s})).
    \end{displaymath}

    \begin{figure}[H]
        \centering
        \includegraphics{../images/embeddImage.pdf}
        \caption{\label{fig:proof:M-estimator-asym-normal-on-manifolds}Depiction of the notation used in the proof of \cref{thm:M-estimator-asym-normal-on-manifolds}. Example with $p = 3$ and $d = \dim(\Theta) = 2$.}
    \end{figure}

    By assumption, the function $M(\mat{\theta})$ is twice continuously differentiable in a neighborhood\footnote{A set $N$ is called a neighborhood of $u$ if there exists an open set $O$ such that $u\in O\subseteq N$.} of $\mat{\theta}_0$. Without loss of generality, we can assume that $U$ is contained in that neighborhood. Then, using the chain rule, we compute the gradient of $M_{\varphi}$ at $\mat{s}_0$ to be $\mat{0}$,
    \begin{displaymath}
        \nabla M_{\varphi}(\mat{s}_0) = {\nabla\varphi^{-1}(\mat{s}_0)}{\nabla M(\varphi^{-1}(\mat{s}_0))} = {\nabla\varphi^{-1}(\mat{s}_0)}{\nabla M(\mat{\theta}_0)} = {\nabla\varphi^{-1}(\mat{s}_0)}\mat{0} = \mat{0}
    \end{displaymath}
    because $\mat{\theta}_0 = \varphi^{-1}(\mat{s}_0)$ is a maximizer of $M$. The second-derivative of $M$, evaluated at $\mat{s}_0 = \varphi(\mat{\theta}_0)$ and using $\nabla M_{\varphi}(\mat{s}_0) = \mat{0}$, is
    \begin{displaymath}
        \nabla^2 M_{\varphi}(\mat{s}_0)
        = \nabla\varphi^{-1}(\mat{s}_0)\nabla^2 M(\varphi^{-1}(\mat{s}_0))\t{\nabla\varphi^{-1}(\mat{s}_0)}
        = \nabla\varphi^{-1}(\mat{s}_0)\mat{H}_{\mat{\theta}_0}\t{\nabla\varphi^{-1}(\mat{s}_0)}.
    \end{displaymath}
    This gives the second-order Taylor expansion of $M_{\varphi}$ at $\mat{s}_0$ as
    \begin{displaymath}
        M_{\varphi}(\mat{s}) = M_{\varphi}(\mat{s}_0) + \frac{1}{2}\t{(\mat{s} - \mat{s}_0)} \nabla^2 M_{\varphi}(\mat{s}_0) (\mat{s} - \mat{s}_0) + \mathcal{O}(\|\mat{s} - \mat{s}_0\|^3)
    \end{displaymath}
    We also need to check the local Lipschitz condition of $m_{\varphi^{-1}(\mat{s})}$. Let $V_{\epsilon}(\mat{s}_0)$ be the open $\epsilon$-ball with center $\mat{s}_0$; i.e.,  $V_{\epsilon}(\mat{s}_0) = \{ \mat{s}\in\mathbb{R}^d : \|\mat{s} - \mat{s}_0\| < \epsilon \}$. Since $\varphi(U)$ contains $\mat{s}_0$, and is open in $\mathbb{R}^d$, there exists an $\epsilon > 0$ such that $V_{\epsilon}(\mat{s}_0)\subseteq\varphi(U)$. The closed $\epsilon/2$ ball $\overline{V}_{\epsilon / 2}(\mat{s}_0)$ is a neighborhood of $\mat{s}_0$ and $\sup_{\mat{s}\in \overline{V}_{\epsilon / 2}(\mat{s}_0)}\|\nabla\varphi^{-1}(\mat{s})\| < \infty$ due to the continuity of $\nabla\varphi^{-1}$ on $\varphi(U)$ with $\overline{V}_{\epsilon / 2}(\mat{s}_0)\subset V_{\epsilon}(\mat{s}_0)\subseteq\varphi(U)$. Then, for almost every $z$ and every $\mat{s}_1 = \varphi(\mat{\theta}_1), \mat{s}_2 = \varphi(\mat{\theta}_2)\in\overline{V}_{\epsilon / 2}(\mat{s}_0)$,
    \begin{multline*}
        | m_{\varphi^{-1}(\mat{s}_1)}(z) - m_{\varphi^{-1}(\mat{s}_2)}(z) |
        = | m_{\mat{\theta}_1}(z) - m_{\mat{\theta}_2}(z) |
        \overset{(a)}{\leq} u(z) \| \mat{\theta}_1 - \mat{\theta}_2 \| \\
        = u(z) \| \varphi^{-1}(\mat{s}_1) - \varphi^{-1}(\mat{s}_2) \|
        \overset{(b)}{\leq} u(z) \sup_{\mat{s}\in \overline{V}_{\epsilon / 2}(\mat{s}_0)}\|\nabla\varphi^{-1}(\mat{s})\| \|\mat{s}_1 - \mat{s}_2\|
        =: v(z) \|\mat{s}_1 - \mat{s}_2\|.
    \end{multline*}
    Here, $(a)$ holds by assumption and $(b)$ is a result of the mean value theorem. Now, $v(z)$ is measurable and square integrable as a scaled version of $u(z)$. Finally, since $\varphi$ is one-to-one, $\hat{\mat{s}}_n = \varphi(\hat{\mat{\theta}}_n)$ is a strong M-estimator for $\mat{s}_0 = \varphi(\mat{\theta}_0)$ of the objective $M_{\varphi}$. We next apply Theorem~5.23 in \cite{vanderVaart1998} to obtain the asymptotic normality of $\hat{\mat{s}}_n$,
    \begin{displaymath}
        \sqrt{n}(\hat{\mat{s}}_n - \mat{s}_0) \xrightarrow{d} \mathcal{N}_{d}(0, \mat{\Sigma}_{\mat{s}_0}),
    \end{displaymath}
    where the $d\times d$ variance-covariance matrix $\mat{\Sigma}_{\mat{s}_0}$ is given by
    \begin{align*}
        \mat{\Sigma}_{\mat{s}_0} &= (\nabla^2 M_{\varphi}(\mat{s}_0))^{-1}\E[\nabla_{\mat{s}} m_{\varphi^{-1}(\mat{s}_0)}(Z)\t{(\nabla_{\mat{s}} m_{\varphi^{-1}(\mat{s}_0)}(Z))}](\nabla^2 M_{\varphi}(\mat{s}_0))^{-1}.
    \end{align*}
    {
        \def\PP{\mat{\varPhi}_{\mat{\theta}_0}}
        \def\EE#1#2{\E[\nabla_{#2} m_{#1}(Z)\t{(\nabla_{#2} m_{#1}(Z))}]}
       Applying the delta method, we obtain
        \begin{displaymath}
            \sqrt{n}(\hat{\mat{\theta}}_n - \mat{\theta}_0)= \sqrt{n}(\varphi^{-1}(\hat{\mat{s}}_n) - \varphi^{-1}(\mat{s}_0))
            \xrightarrow{d} \mathcal{N}_p(0, \t{\nabla\varphi^{-1}(\mat{s}_0)}\mat{\Sigma}_{\mat{s}_0}{\nabla\varphi^{-1}(\mat{s}_0)}).
        \end{displaymath}
        We continue by reexpressing the $p\times p$ asymptotic variance-covariance matrix of $\hat{\mat{\theta}}_n$ in terms of $\mat{\theta}_0$ instead of $\mat{s}_0 = \varphi(\mat{\theta}_0)$. We let $\PP = \t{\nabla\varphi^{-1}(\varphi(\mat{\theta}_0))} = \t{\nabla\varphi^{-1}(\mat{s}_0)}$ and observe that for all $\mat{s}\in\varphi(U)$, the gradient of $\mat{s}\mapsto m_{\varphi^{-1}(\mat{s})}(z)$ evaluated at $\mat{s}_0 = \varphi(\mat{\theta}_0)$ has the form
        \begin{displaymath}
            \nabla_{\mat{s}}m_{\varphi^{-1}(\mat{s}_0)}(z)= \nabla\varphi^{-1}(\mat{s}_0)\nabla_{\mat{\theta}}m_{\mat{\theta}_0}(z)= \t{\PP}\nabla_{\mat{\theta}}m_{\mat{\theta}_0}(z).
        \end{displaymath}
        Then,
        \begin{multline*}
            \t{\nabla\varphi^{-1}(\mat{s}_0)}\mat{\Sigma}_{\mat{s}_0}{\nabla\varphi^{-1}(\mat{s}_0)}= \PP\mat{\Sigma}_{\mat{s}_0}\t{\PP} \\
            = \PP(\nabla^2 M_{\varphi}(\mat{s}_0))^{-1}\EE{\varphi^{-1}(\mat{s}_0)}{}(\nabla^2 M_{\varphi}(\mat{s}_0))^{-1}\t{\PP} \\
            = {\PP}(\t{\PP}\mat{H}_{\mat{\theta}_0}\PP)^{-1}\t{\PP}\EE{\mat{\theta}_0}{\mat{\theta}}{\PP}(\t{\PP}\mat{H}_{\mat{\theta}_0}\PP)^{-1}\t{\PP} \\
            = \mat{\Pi}_{\mat{\theta}_0}\EE{\mat{\theta}_0}{\mat{\theta}}\mat{\Pi}_{\mat{\theta}_0}
        \end{multline*}
        where the last equality holds because $\Span\PP = T_{\mat{\theta}_0}\Theta$, by \cref{def:tangent-space} of the tangent space $T_{\mat{\theta}_0}\Theta$.

        It remains to show that $\mat{\Pi}_{\mat{\theta}_0} = \mat{P}_{\mat{\theta}_0}\pinv{(\t{\mat{P}_{\mat{\theta}_0}}\mat{H}_{\mat{\theta}_0}\mat{P}_{\mat{\theta}_0})}\t{\mat{P}_{\mat{\theta}_0}}$ for any $p\times k$ matrix $\mat{P}_{\mat{\theta}_0}$ such that $k\geq d$ and $\Span{\mat{P}_{\mat{\theta}_0}} = T_{\mat{\theta}_0}\Theta$. This also ensures that the final result is independent of the chosen chart $\varphi$, since the tangent space does not depend on a specific chart. Therefore, let $\PP = {\mat{Q}}{\mat{R}}$ and $\mat{P}_{\mat{\theta}_0} = \widetilde{\mat{Q}}\widetilde{\mat{R}}$ be their thin QR decompositions, respectively. Both $\mat{Q}, \widetilde{\mat{Q}}$ have dimension $p\times d$,   $\mat{Q}$ is semi-orthogonal, and $\mat{R}$ is invertible of dimension $d\times d$ while $\widetilde{\mat{R}}$ is a $d\times k$ full row-rank matrix. Since $\mat{Q}$ is semi-orthogonal, the $p\times p$ matrix $\mat{Q}\t{\mat{Q}}$ is an orthogonal projection onto $\Span\mat{Q} = \Span\mat{P}_{\mat{\theta}_0} = T_{\mat{\theta}_0}\Theta$. This allows to express $\mat{P}_{\mat{\theta}_0}$ in terms of $\mat{Q}$ as
        \begin{displaymath}
            \mat{P}_{\mat{\theta}_0} = \mat{Q}\t{\mat{Q}}\mat{P}_{\mat{\theta}_0}
            = \mat{Q}\t{\mat{Q}}\widetilde{\mat{Q}}\widetilde{\mat{R}} =: {\mat{Q}}\mat{M}.
        \end{displaymath}
        From $\Span\mat{Q} = \Span\mat{P}_{\mat{\theta}_0}$, it follows that the $d\times k$ matrix $\mat{M}$ is also of full row-rank. Also, $\mat{M}\pinv{\mat{M}} = \mat{I}_d = \mat{R}\mat{R}^{-1}$ as a property of the Moore-Penrose pseudo inverse with $\mat{M}$ being of full row-rank. Another property of the pseudo inverse is that for matrices $\mat{A}, \mat{B}$, where $\mat{A}$ has full column-rank and $\mat{B}$ has full row-rank, $\pinv{(\mat{A}\mat{B})} = \pinv{\mat{B}}\pinv{\mat{A}}$. This enables the computation
        \begin{multline*}
            \mat{P}_{\mat{\theta}_0}\pinv{(\t{\mat{P}_{\mat{\theta}_0}}\mat{H}_{\mat{\theta}_0}\mat{P}_{\mat{\theta}_0})}\t{\mat{P}_{\mat{\theta}_0}}= \mat{Q} \mat{M} \pinv{\mat{M}} (\t{\mat{Q}} \mat{H}_{\mat{\theta}_0} \mat{Q})^{-1} \t{(\mat{M} \pinv{\mat{M}})} \t{\mat{Q}} \\
            = \mat{Q} {\mat{R}} {\mat{R}}^{-1} (\t{\mat{Q}} \mat{H}_{\mat{\theta}_0} \mat{Q})^{-1} \t{({\mat{R}} {\mat{R}}^{-1})} \t{\mat{Q}}
            = \PP(\t{\PP}\mat{H}_{\mat{\theta}_0}\PP)^{-1}\t{\PP}
            = \mat{\Pi}_{\mat{\theta}_0}.
        \end{multline*}
    }
\end{proof}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Asymptotic Normality of GMLM}
We rewrite the log-likelihood \eqref{eq:log-likelihood} in a different form to simplify the proof of \cref{thm:asymptotic-normality-gmlm} and to provide the notation to express the regularity conditions of \cref{thm:asymptotic-normality-gmlm} in a compact form.

The first natural parameter component $\mat{\eta}_{1y}$ defined in \eqref{eq:eta1-manifold} can be written as
\begin{align*}
    \mat{\eta}_{1y}
    &= \vec{\overline{\ten{\eta}}} + \mat{B}\vec{\ten{F}_y}= \mat{I}_p\vec{\overline{\ten{\eta}}} + (\t{(\vec{\ten{F}_y})}\otimes\mat{I}_p)\vec{\mat{B}} \\
    &= \begin{pmatrix}
        \mat{I}_p & \t{(\vec{\ten{F}_y})}\otimes\mat{I}_p
    \end{pmatrix}\begin{pmatrix}
        \vec{\overline{\ten{\eta}}} \\
        \vec{\mat{B}}
    \end{pmatrix}.
\end{align*}
The second natural parameter component $\mat{\eta}_2$, modeled in \eqref{eq:eta2-manifold}, relates to $\vech{\mat{\Omega}}$ linearly as
\begin{displaymath}
    \mat{\eta}_2 = c\t{\mat{D}_p}\vec{\mat{\Omega}} = c\t{\mat{D}_p}\mat{D}_p\vech{\mat{\Omega}}.
\end{displaymath}
This gives the following relation between $\mat{\eta}_y = (\mat{\eta}_{1y}, \mat{\eta}_2)$ and $\mat{\xi} = (\vec{\overline{\ten{\eta}}}, \vec{\mat{B}}, \vech{\mat{\Omega}})\in\Xi$,
\begin{equation}
    \mat{\eta}_y = \begin{pmatrix}
        \mat{I}_p & \ \ \ \t{(\vec{\ten{F}_y})}\otimes\mat{I}_p\ \ & 0 \\
        0 & 0 & c\t{\mat{D}_p}\mat{D}_p
    \end{pmatrix}
\begin{pmatrix}
\vec{\overline{\ten{\eta}}} \\
\vec{\mat{B}} \\
\vech{\mat{\Omega}}
\end{pmatrix} =: \mat{F}(y)\mat{\xi} \label{eq:eta-to-xi-linear-relation}
\end{equation}
where $\mat{F}(y)$ is a $p (p + 3) / 2\times p (p + 2 q + 3) / 2$ dimensional matrix-valued function in $y$. Moreover, for every $y$ the matrix $\mat{F}(y)$ is of full rank.
The log-likelihood of model \eqref{eq:quad-density} for the unconstrained parameters $\xi\in\Xi$ is
\begin{displaymath}
    l_n(\mat{\xi})
        = \frac{1}{n}\sum_{i = 1}^{n} (\langle \mat{t}(\ten{X}), \mat{\eta}_{y} \rangle - b(\mat{\eta}_y))
        =: \frac{1}{n}\sum_{i = 1}^{n} m_{\mat{\xi}}(Z_i)
\end{displaymath}
where $Z_i = (\ten{X}_i, Y_i)$. Using \eqref{eq:eta-to-xi-linear-relation} we can write
\begin{displaymath}
    m_{\mat{\xi}}(z) = \langle\mat{t}(\ten{X}), \mat{F}(y)\mat{\xi}\rangle - b(\mat{F}(y)\mat{\xi}).
\end{displaymath}

The following are the regularity conditions for the log-likelihood required by \cref{thm:asymptotic-normality-gmlm}.

\begin{condition}\label{cond:differentiable-and-convex}
The mapping $\mat{\xi}\mapsto m_{\mat{\xi}}(z)$ is twice continuously differentiable for almost every $z$ and $z\mapsto m_{\mat{\xi}}(z)$ is measurable. Moreover, $\mat{\eta}\mapsto b(\mat{\eta})$ is strictly convex. Furthermore, for every $\widetilde{\mat{\eta}}$, $P(\mat{F}(Y)\mat{\xi} = \widetilde{\mat{\eta}}) < 1$.
\end{condition}

\begin{condition}\label{cond:moments}
$\E\|\t{\mat{t}(\ten{X})}\mat{F}(Y)\| < \infty$,  and $\E\|\t{\mat{t}(\ten{X})}\mat{F}(Y)\|^2 < \infty$.
\end{condition}

\begin{condition}\label{cond:finite-sup-on-compacta}
The mapping $\mat{\eta}\mapsto b(\mat{\eta})$ is twice continuously differentiable and for every non-empty compact $K\subseteq\Xi$, \begin{gather*}
\E\sup_{\mat{\xi}\in K}\|b(\mat{F}(Y)\mat{\xi})\| < \infty, \qquad
\E\sup_{\mat{\xi}\in K}\|\t{\nabla b(\mat{F}(Y)\mat{\xi})}\mat{F}(Y)\|^2 < \infty, \\
\E\sup_{\mat{\xi}\in K}\| \t{\mat{F}(Y)}\nabla^2 b(\mat{F}(Y)\mat{\xi})\mat{F}(Y) \| < \infty.
\end{gather*}
\end{condition}

We continue with some more technical lemmas needed for the proof of \cref{thm:asymptotic-normality-gmlm}.

\begin{lemma}\label{thm:kron-perm}
    Given $r \geq 2$ matrices $\mat{A}_k$ of dimension $p_j\times q_j$ for $k = 1, \ldots, r$, then there exists a unique permutation matrix $\mat{S}_{\mat{p}, \mat{q}}$ such that
    \begin{equation}\label{eq:kron-to-outer-perm}
        \vec\bigkron_{k = r}^{1}\mat{A}_k = \mat{S}_{\mat{p}, \mat{q}}\vec\bigouter_{k = 1}^{r}\mat{A}_k.
    \end{equation}
    The permutation $\mat{S}_{\mat{p}, \mat{q}}$ with indices $\mat{p} = (p_1, \ldots, p_r)$ and $\mat{q} = (q_1, \ldots, q_r)$ is defined recursively as
    \begin{equation}\label{eq:S_pq}
        \mat{S}_{\mat{p}, \mat{q}} = \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)} \bigl(\mat{I}_{p_r q_r}\otimes\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\bigr)
    \end{equation}
    with initial value
    \begin{displaymath}
        \mat{S}_{(p_1, p_2), (q_1, q_2)} = \mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1}
    \end{displaymath}
    where $\mat{K}_{p, q}$ is the \emph{commutation matrix} \cite[][Ch.~11]{AbadirMagnus2005};  that is, the permutation such that $\vec{\t{\mat{A}}} = \mat{K}_{p, q}\vec{\mat{A}}$ for every $p\times q$ dimensional matrix $\mat{A}$.
\end{lemma}
\begin{proof}
    Lemma~7 in \cite{MagnusNeudecker1986} states that
    \begin{align}
    \vec(\mat{A}_2\otimes\mat{A}_1)
    &= (\mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1})(\vec{\mat{A}_2}\otimes\vec{\mat{A}_1}) \label{eq:MagnusNeudecker1986-vec-kron-identity} \\
    &= (\mat{I}_{q_2}\otimes\mat{K}_{q_1, p_2}\otimes\mat{I}_{p_1})\vec(\mat{A}_1\circ \mat{A}_2). \nonumber
    \end{align}
    This proves the statement for $r = 2$. The general statement for $r > 2$ follows by induction. Assuming \eqref{eq:kron-to-outer-perm} holds for $r - 1$, the induction step is 
    \begin{multline*}
    \vec{\bigkron_{k = r}^{1}}\mat{A}_k
    = \vec\Bigl(\mat{A}_r\otimes\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr) \\
    \overset{\eqref{eq:MagnusNeudecker1986-vec-kron-identity}}{=} \Bigl( \mat{I}_{q_r}\otimes\mat{K}_{\prod_{k = 1}^{r - 1}q_k, p_r}\otimes\mat{I}_{\prod_{k = 1}^{r - 1}p_k} \Bigr)\vec\Bigl((\vec\mat{A}_r)\otimes\vec\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr) \\
    = \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)}\vec\Bigl[\Bigl(\vec\bigkron_{k = r - 1}^{1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
    \overset{\eqref{eq:kron-to-outer-perm}}{=} \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)}\vec\Bigl[\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\Bigl(\vec\bigouter_{k = 1}^{r - 1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
    \overset{(a)}{=} \mat{S}_{\bigl( \prod_{k = 1}^{r - 1}p_k, p_r \bigr), \bigl( \prod_{k = 1}^{r - 1}q_k, q_r \bigr)} \bigl(\mat{I}_{p_rq_r}\otimes\mat{S}_{(p_1, \ldots, p_{r-1}), (q_1, \ldots, q_{r-1})}\bigr)\vec\Bigl[\Bigl(\vec\bigouter_{k = 1}^{r - 1}\mat{A}_k\Bigr)\t{(\vec\mat{A}_r)}\Bigr] \\
    =\mat{S}_{\mat{p},\mat{q}}\vec\bigouter_{k = 1}^{r}\mat{A}_k.
    \end{multline*}
    Equality $(a)$ uses the relation $\vec(\mat{C}\mat{a}\t{\mat{b}}) = (\mat{I}_{\dim(\mat{b})}\otimes\mat{C})\vec(\mat{a}\t{\mat{b}})$ for a matrix $\mat{C}$ and vectors $\mat{a}, \mat{b}$.
\end{proof}

\begin{lemma}\label{thm:kron-manifold-tangent-space}
    Let $\manifold{A}_k\subseteq\mathbb{R}^{p_k\times q_k}\backslash\{\mat{0}\}$ for $k = 1, \ldots, r$ be smooth embedded submanifolds as well as ether a sphere or a cone. Then
    \begin{displaymath}
        \manifold{K} = \Bigl\{ \bigkron_{k = r}^{1}\mat{A}_k : \mat{A}_k\in\manifold{A}_k \Bigr\}
    \end{displaymath}
    is an embedded manifold in $\mathbb{R}^{p\times q}$ for $p = \prod_{k = 1}^{r} p_k$ and $q = \prod_{k = 1}^{r} q_k$.
    Furthermore, define for $j = 1, \ldots, r$ the matrices
    \begin{equation}\label{eq:kron-differential-span}
        \mat{\Gamma}_j
            = \bigkron_{k = r}^{1}(\mat{I}_{p_k q_k}\mathrm{\ if\ } j = k \mathrm{\ else\ }\vec{\mat{A}_k})
            = \bigkron_{k = r}^{j + 1}(\vec{\mat{A}_k})\otimes\mat{I}_{p_j q_j}\otimes\bigkron_{k = j - 1}^{1}(\vec{\mat{A}_k})
    \end{equation}
    and let $\gamma_j$ be $p_j q_j\times d_j$ matrices with $d_j \geq\dim\manifold{A}_j$ which span the tangent space $T_{\mat{A}_j}\manifold{A}_j$ of $\manifold{A}$ at $\mat{A}_j\in\manifold{A}_j$, that is $\Span\gamma_j = T_{\mat{A}_j}\manifold{A}_j$.
    Then, with the permutation matrix $\mat{S}_{\mat{p}, \mat{q}}$ defined in \eqref{eq:S_pq}, the $p q \times \sum_{k = 1}^{r} d_j$ dimensional matrix
    \begin{displaymath}
        \mat{P}_{\mat{A}} = \mat{S}_{\mat{p}, \mat{q}}\left[\mat{\Gamma}_1\mat{\gamma}_1, \mat{\Gamma}_2\mat{\gamma}_2, \ldots, \mat{\Gamma}_r\mat{\gamma}_r\right]
    \end{displaymath}
    spans the tangent space $T_{\mat{A}}\manifold{K}$ of $\manifold{K}$ at $\mat{A} = \bigkron_{k = r}^{1}\mat{A}_k\in\manifold{K}$, in formula $\Span\mat{P}_{\mat{A}} = T_{\mat{A}}\manifold{K}$.
\end{lemma}
\begin{proof}
    The statement that $\manifold{K}$ is an embedded manifold follows via induction using \cref{thm:kron-manifolds}.
    We compute the differential of the vectorized Kronecker product using \cref{thm:kron-perm} where $\mat{S}_{\mat{p}, \mat{q}}$ is the permutation \eqref{eq:S_pq} defined therein.
    \begin{multline*}
        \d\vec\bigotimes_{k = r}^{1}\mat{A}_k
        = \vec\sum_{j = 1}^{r}\bigkron_{k = r}^{1}(\ternary{k = j}{\d\mat{A}_j}{\mat{A}_k}) \\
        = \mat{S}_{\mat{p},\mat{q}}\vec\sum_{j = 1}^{r}\Bigl(\bigouter_{k = 1}^{r}(\ternary{k = j}{\d\mat{A}_j}{\mat{A}_k})\Bigr) \\
        = \mat{S}_{\mat{p}, \mat{q}}\sum_{j = 1}^{r}\bigkron_{k = r}^{1}(\ternary{k = j}{\vec\d\mat{A}_j}{\vec\mat{A}_k}) \\
        =\mat{S}_{\mat{p}, \mat{q}}\sum_{j =1}^{r}\Bigl(\bigkron_{k = r}^{1}(\ternary{k = j}{\mat{I}_{p_j q_j}}{\vec\mat{A}_k})\Bigr)\vec\d\mat{A}_j \\
        = \mat{S}_{\mat{p}, \mat{q}}\sum_{j = 1}^{r}\mat{\Gamma}_j\vec\d\mat{A}_j
        = \mat{S}_{\mat{p}, \mat{q}}[\mat{\Gamma}_1, \ldots, \mat{\Gamma}_r]\begin{pmatrix}
            \vec\d\mat{A}_1 \\ \vdots \\ \vec\d\mat{A}_r
        \end{pmatrix}
    \end{multline*}
    Due to the definition of the manifold, this differential provides the gradient of a surjective map into the manifold. The span of the gradient then spans the tangent space.

    Now, we take a closer look at the differentials $\vec{\d\mat{A}_j}$ for $j = 1, \ldots, r$. Let $\varphi_j$ be a chart of $\manifold{A}_j$ in a neighborhood of $\mat{A}_j$. Then, $\mat{A}_j = \varphi_j^{-1}(\varphi_j(\mat{A}_j))$ which gives
    \begin{displaymath}\vec{\d\mat{A}_j} = \t{\nabla\varphi_j^{-1}(\varphi_j(\mat{A}_j))}\vec\d\varphi_j(\mat{A}_j).
    \end{displaymath}
    Therefore, for every matrix $\mat{\gamma}_j$ such that $\Span{\mat{\gamma}_j} = T_{\mat{A}_j}\manifold{A}_j$ holds $\Span{\t{\nabla\varphi_j^{-1}(\varphi_j(\mat{A}_j))}} = \Span{\mat{\gamma}_j}$ by \cref{def:tangent-space} of the tangent space. We get
    \begin{displaymath}
        \Span\mat{S}_{\mat{p}, \mat{q}}[\mat{\Gamma}_1, \ldots, \mat{\Gamma}_r]\begin{pmatrix}
            \vec\d\mat{A}_1 \\ \vdots \\ \vec\d\mat{A}_r
        \end{pmatrix}
        =\Span\mat{S}_{\mat{p}, \mat{q}}[\mat{\Gamma}_1\mat{\gamma}_1, \ldots, \mat{\Gamma}_r\mat{\gamma}_r]
        =\Span\mat{P}_{\mat{A}}
    \end{displaymath}
    which concludes the proof.
\end{proof}

\begin{proof}[\hypertarget{proof:asymptotic-normality-gmlm}{Proof of \cref{thm:asymptotic-normality-gmlm}}]
    The proof consists of three parts. First, we show the existence of a consistent strong M-estimator by applying \cref{thm:M-estimator-consistency-on-subsets}. Next, we apply \cref{thm:M-estimator-asym-normal-on-manifolds} to obtain its asymptotic normality. We conclude by computing the missing parts of the asymtotic covariance matrix $\mat{\Sigma}_{\mat{\theta}_0}$ provided by \cref{thm:M-estimator-asym-normal-on-manifolds}.

    We check whether the conditions of \cref{thm:M-estimator-consistency-on-subsets} are satisfied. On $\Xi$, the mapping $\mat{\xi}\mapsto m_{\mat{\xi}}(z) = m_{\mat{\xi}}(\ten{X},y) = \langle \mat{F}(y)\mat{\xi}, \mat{t}(\ten{X}) \rangle - b(\mat{F}(y)\mat{\xi})$ is strictly concave for every $z$ because $\mat{\xi}\mapsto\mat{F}(y)\mat{\xi}$ is linear and $b$ is strictly convex by \cref{cond:differentiable-and-convex}. Since $\ten{X} \mid Y$ is distributed according to \eqref{eq:quad-density}, the function $M(\mat{\xi}) = \E m_{\mat{\xi}}(Z)$ is well defined by \cref{cond:moments}. Let $\mat{\xi}_k = (\vec{\overline{\ten{\eta}}_k}, \vec{\mat{B}_k}, \vech{\mat{\Omega}_k})$, and $f_{\mat{\xi}_k}$ be the pdf of $\ten{X} \mid Y$ indexed by $\mat{\xi}_k$, for $k = 1, 2$. If $\mat{\xi}_1\ne \mat{\xi}_2$, then $f_{\mat{\xi}_1} \neq f_{\mat{\xi}_2}$, which obtains that the true $\mat{\theta}_0$ is a unique maximizer of $\mat{\theta}_0\in\Theta\subseteq\Xi$ by applying Lemma~5.35 from \cite{vanderVaart1998}. Finally, under \cref{cond:finite-sup-on-compacta}, all assumptions of \cref{thm:M-estimator-consistency-on-subsets} are fulfilled yielding the existence of a consistent strong M-estimator over $\Theta\subseteq\Xi$.

    Next, let $\hat{\mat{\theta}}_n$ be a strong M-estimator on $\Theta\subseteq\Xi$, whose existence and consistency was shown in the previous step. Since $z\mapsto m_{\mat{\xi}}(z)$ is measurable for all $\mat{\xi}\in\Xi$, it is also measurable in a neighborhood of $\mat{\theta}_0$. The differentiability of  $\mat{\theta}\mapsto m_{\mat{\theta}}(z)$ is stated in \cref{cond:differentiable-and-convex}. For the Lipschitz condition, let $K\subseteq\Xi$ be a compact neighborhood of $\mat{\theta}_0$, which exists since $\Xi$ is open. Then,
    \begin{multline*}
        \left| m_{\mat{\theta}_1}(z) - m_{\mat{\theta}_2}(z) \right|
        = \left| \langle \mat{t}(\ten{X}), \mat{F}(y)(\mat{\theta}_1 - \mat{\theta}_2) \rangle - b(\mat{F}(z)\mat{\theta}_1) + b(\mat{F}(z)\mat{\theta}_2) \right| \\
        \leq (\| \t{\mat{F}(y)}\mat{t}(\ten{X}) \|_2 + \sup_{\mat{\theta}\in K}\| \nabla b(\mat{F}(y)\mat{\theta}) \mat{F}(y)\| ) \| \mat{\theta}_1 - \mat{\theta}_2 \|_2
        =: u(z)\| \mat{\theta}_1 - \mat{\theta}_2 \|_2
    \end{multline*}
    with $u(z)$ being measurable and square integrable derives from \cref{cond:finite-sup-on-compacta}. The existence of a second-order Taylor expansion of $\mat{\theta}\mapsto M(\mat{\theta}) = \E m_{\mat{\theta}}(Z)$ in a neighborhood of $\mat{\theta}_0$ holds by \cref{cond:finite-sup-on-compacta}. Moreover, the Hessian $\mat{H}_{\mat{\theta}_0}$ is non-singular by the strict convexity of $b$ stated in \cref{cond:differentiable-and-convex}. Now, we can apply \cref{thm:M-estimator-asym-normal-on-manifolds} to obtain the asymptotic normality of $\sqrt{n}(\hat{\mat{\theta}}_n - \mat{\theta}_0)$ with variance-covariance structure 
    \begin{equation}\label{eq:asymptotic-covariance-gmlm}
        \mat{\Sigma}_{\mat{\theta}_0} = \mat{\Pi}_{\mat{\theta}_0} \E[\nabla m_{\mat{\theta}_0}(Z)\t{(\nabla m_{\mat{\theta}_0}(Z))}]\mat{\Pi}_{\mat{\theta}_0}
    \end{equation}
    where $\mat{\Pi}_{\mat{\theta}_0} = \mat{P}_{\mat{\theta}_0}(\t{\mat{P}_{\mat{\theta}_0}}\mat{H}_{\mat{\theta}_0}\mat{P}_{\mat{\theta}_0})^{-1}\t{\mat{P}_{\mat{\theta}_0}}$ and $\mat{P}_{\mat{\theta}_0}$ is any $p\times \dim(\Theta)$ matrix such that it spans the tangent space of $\Theta$ at $\mat{\theta}_0$. That is, $\Span \mat{P}_{\mat{\theta}_0} = T_{\mat{\theta}_0}\Theta$.

    Finally, we compute a matrix $\mat{P}_{\mat{\theta}_0}$ such that $\Span{\mat{P}_{\mat{\theta}_0}} = T_{\mat{\theta}_0}\Theta$ for $\Theta = \mathbb{R}^p\times\manifold{K}_{\mat{B}}\times\vech(\manifold{K}_{\mat{\Omega}})$ as in \cref{thm:param-manifold}. Since the manifold $\Theta$ is a product manifold we get a block diagonal structure for $\mat{P}_{\mat{\theta}_0}$ as
    \begin{displaymath}
        \mat{P}_{\mat{\theta}_0} = \begin{pmatrix}
            \mat{I}_p & 0 & 0 \\
            0 & \mat{P}_{\mat{B}_0} & 0 \\
            0 & 0 & \mat{P}_{\mat{\Omega}_0}
        \end{pmatrix}
    \end{displaymath}
    where $\mat{I}_p$ is the identity matrix spanning the tangent space of $\mathbb{R}^p$, which is identified with $\mathbb{R}^p$ itself. The blocks $\mat{P}_{\mat{B}_0}$ and $\mat{P}_{\mat{\Omega}_0}$ need to span the tangent spaces of $\manifold{K}_{\mat{B}}$ and $\vech(\manifold{K}_{\mat{\Omega}})$, respectively. Both $\manifold{K}_{\mat{B}}$ and $\vech(\manifold{K}_{\mat{\Omega}})$ are manifolds according to \cref{thm:param-manifold}. The manifold $\vech(\manifold{K}_{\mat{\Omega}})$ is the image of $\manifold{K}_{\mat{\Omega}}$ under the linear map $\vech(\mat{\Omega}) = \pinv{\mat{D}_p}\vec{\mat{\Omega}}$ leading to the differential $\d\vech{\mat{\Omega}} = \pinv{\mat{D}_p}\vec{\d\mat{\Omega}}$. By applying \cref{thm:kron-manifold-tangent-space} for $\manifold{K}_{\mat{B}}$ and $\manifold{K}_{\mat{\Omega}}$ gives    
    \begin{align*}
        \mat{P}_{\mat{B}_0} &= \mat{S}_{\mat{p}, \mat{q}}[\mat{\Gamma}_{\mat{\beta}_1}\mat{\gamma}_{\mat{\beta}_1}, \ldots, \mat{\Gamma}_{\mat{\beta}_r}\mat{\gamma}_{\mat{\beta}_r}], \\
        \mat{P}_{\mat{\Omega}_0} &= \pinv{\mat{D}_p}\mat{S}_{\mat{p}, \mat{p}}[\mat{\Gamma}_{\mat{\Omega}_1}\mat{\gamma}_{\mat{\Omega}_1}, \ldots, \mat{\Gamma}_{\mat{\Omega}_r}\mat{\gamma}_{\mat{\Omega}_r}]
    \end{align*}
    where the matrices $\mat{S}_{\mat{p}, \mat{q}}$, $\mat{\Gamma}_{\mat{\beta}_j}$, $\mat{\gamma}_{\mat{\beta}_j}$, $\mat{\Gamma}_{\mat{\Omega}_j}$ and $\mat{\gamma}_{\mat{\Omega}_j}$ are described in \cref{thm:kron-manifold-tangent-space} for the Kronecker manifolds $\manifold{K}_{\mat{B}}$ and $\manifold{K}_{\mat{\Omega}}$. Leading to
    \begin{equation}\label{eq:param-manifold-span}
        \mat{P}_{\mat{\theta}_0} = \begin{pmatrix}
            \mat{I}_p & 0 & 0 \\
            0 & \mat{S}_{\mat{p}, \mat{q}}[\mat{\Gamma}_{\mat{\beta}_1}\mat{\gamma}_{\mat{\beta}_1}, \ldots, \mat{\Gamma}_{\mat{\beta}_r}\mat{\gamma}_{\mat{\beta}_r}] & 0 \\
            0 & 0 & \pinv{\mat{D}_p}\mat{S}_{\mat{p}, \mat{p}}[\mat{\Gamma}_{\mat{\Omega}_1}\mat{\gamma}_{\mat{\Omega}_1}, \ldots, \mat{\Gamma}_{\mat{\Omega}_r}\mat{\gamma}_{\mat{\Omega}_r}]
        \end{pmatrix}.
    \end{equation}
\end{proof}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Demonstration of the Advantages of Tensor Representations}\label{sec:GMLMvsVecEfficiency}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

%%% Minor revision reviewer comment
% My concerns have been largely resolved. One minor point remains:
%
% I appreciate the authors’ clarification that the Kronecker-structured
% parameterization improves efficiency by exploiting the tensor structure.
% While the rationale is sound, readers would benefit from a brief numerical
% illustration--perhaps in the supplementary material--demonstrating empirical
% variance reduction or faster convergence relative to an unconstrained fit.
% Including a small, illustrative example would make the claimed efficiency gain
% more concrete and convincing.

We illustrate the competitive advantage of tensor structure as compared with vector representation using a simple example. We generate standard normal responses $y_i$, for $i = 1, \ldots, n$.
For this simulation, we set the GMLM parameters $\overline{\ten{\eta}} = \mat{0}$, $\mat{\beta}_1 = \mat{\beta}_2 = \t{(-1, 1, -1, 1, \ldots)}\in\mathbb{R}^{\tilde{p}}$ to vectors of alternating signs, and $\mat{\Sigma}_1 = \mat{\Sigma}_2 = \mat{I}_{\tilde{p}}$ so that $\mat{\alpha}_j=\mat{\Sigma}_j^{-1}\mat{\beta}_j=\mat{\beta}_j$, $j=1,2$, and %the precision matrices 
$\mat{\Omega}_1 = \mat{\Omega}_2 = \mat{I}_{\tilde{p}}$. Given $y_i$, we draw matrix-valued $\ten{X}_i$'s of dimension $p_1\times p_2$, where $p_1 = p_2 = \tilde{p}$ from 
\begin{equation}\label{eq:bilinear}
    \ten{X}_i = \bar{\mat{\eta}} + \mat{\alpha}_1\ten{F}_{y_i}\t{\mat{\alpha}_2} + \mat{\epsilon}_i
\end{equation}
where $\ten{F}_{y_i}\equiv y_i$ is a $1\times 1$ matrix, $\vec{\mat{\epsilon}_i}\sim\mathcal{N}(\mat{0}, \mat{\Sigma}_2\otimes\mat{\Sigma}_1)$. That is,  $\ten{X}\mid Y = y_i$ is matrix normal with parameters as specified above.
 

We compare four different estimators for the vectorized reduction matrix $\mat{B} = \mat{\beta}_2\otimes\mat{\beta}_1\in\mathbb{R}^{\tilde{p}^2}$. First, a linear model, which, in this scenario, is equivalent to the vectorized GMLM model if the sample size exceeds the dimensionality $\tilde{p}^2$ of the vectorized predictors $\vec(\ten{X})$. Second, our GMLM model applied to the vectorized predictors, which are distributed as multi-variate normal (tensor order $r = 1$) conditionally on $y$. Third, our GMLM model as designed for matrix-valued predictors (tensor order $r = 2$). Fourth, another GMLM model for matrix-valued predictors but now with a manifold constraint on the precision matrices $\mat{\Omega}_1, \mat{\Omega}_2$ to be scaled diagonal matrices. This means we let $\mat{\Omega}_1, \mat{\Omega}_2\in\{ a\mat{I}_{\tilde{p}} : 0 < a \}$, which is a one-dimensional matrix manifold from \cref{tab:matrix-manifolds}. We did not specify $a$ to be 1 to operate in a context where there is not enough information to uniquely resolve the scale. In the first two cases the reduction matrix estimate, $\hat{\mat{B}}$, is computed directly. In contrast, the matrix-valued GMLM models provide the mode-wise estimates $\hat{\mat{\beta}}_1, \hat{\mat{\beta}}_2$ which are then used to compute $\hat{\mat{B}} = \hat{\mat{\beta}}_2\otimes\hat{\mat{\beta}}_1$.

For accuracy assessment we use the subspace distance $d(\mat{B}, \hat{\mat{B}})$ from \cref{sec:simulations} for comparison of the estimates $\hat{\mat{B}}$ versus the true reduction $\mat{B} = \mat{\beta}_2\otimes\mat{\beta}_1$ used for data generation.

The simulation was carried out for different sample sizes $n = 100, 200, 300, 500, 750$ and increasing matrix row and column dimension $\tilde{p} = 2, 3, 4, 5, 6, 7, 9, 11, 13, 15, 18, 22, 27, 32$ of the predictors $\ten{X}$. The results are visualized in \cref{fig:GMLMvsVecEfficiency}, where we plot the mean $\pm$ one standard deviation of the subspace distance over 100 replicates. 

Both GMLM estimators perform uniformly better irrespective of $\tilde{p}$ that ranges from 2 to 32 in this simulation. Between these two, the constrained model, which is the closest to the true variance-covariance structure,  practically reproduces the true model. Also, they both exhibit roughly the same accuracy, and in particular no deterioration, despite the increase in dimension $\tilde{p}$.

The linear model performs as expected. It drastically deteriorates as the dimension increases. The yellow lines stop where the dimension $\tilde{p}^2$ exceeds the sample size $n$. Similarly, the GMLM applied to the vectorized matrix-valued $\ten{X}$ has identical performance as the plain vanilla linear model up to the point that the regularization kicks in, as described in \cref{sec:tensor-normal-estimation}. For example, for $n=750$, the performance of the two models is identical up to $\tilde{p}=18$ and then it improves till it hits its floor. The latter will nevertheless always exceed the corresponding accuracy of either tensor versions of the GMLM model,  with and without the constraint.  

The accuracy is monotone increasing as a function of the sample size across all four models. That is,  the closer the lines are to the $x$-axis the higher the sample size is across $\tilde{p}$. The only exception can be seen in the GMLM applied to the vectorized predictors (red lines) in the transition phase of GMLM using regularization to deal with $\tilde{p}^2 > n$. 

\begin{figure}
    \centering
    \caption{\label{fig:GMLMvsVecEfficiency}Accuracy Gain: Tensor-valued versus Vectorized Predictors}
    \includegraphics[width = \textwidth]{plots/sim_efficiency.pdf}
\end{figure}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Investigation and Comparison of GMLM versus TSIR}\label{sec:GMLMvsTSIR}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

TSIR suffers substantial loss in predictive accuracy in the full 2D EEG data (see \cref{tab:eeg}) as compared to the lower dimensional setting of pre-processed $\ten{X}$. Compelling evidence suggests that this is due to TSIR not directly estimating the mode-wise reduction matrices $\mat{\beta}_j$: It estimates the mode-wise covariances $\mat{\Sigma}_j$ and mode-wise reduction matrices $\mat{\Gamma}_j$ for standardized observations and sets $\widehat{\mat{\beta}}_j = \widehat{\mat{\Sigma}}_j^{-1}\widehat{\mat{\Gamma}}_j$.   The estimation accuracy of the $\widehat{\mat{\beta}}_j$s depends on both $\widehat{\mat{\Sigma}}_j$ and  $\widehat{\mat{\Gamma}}_j$. Specifically, as the dimensions increase, the mode-wise covariances are ill-conditioned,  leading to massive errors in the estimation accuracy of the reduction matrices $\widehat{\mat{\beta}}_j$.

\begin{figure}
    \centering
    \caption{\label{fig:TSIRvsGMLM}TSIR in red vs. GMLM in black}
    \includegraphics[width = \textwidth]{plots/tsir_eeg_2d_bad_perf.pdf}
\end{figure}

To observe this in practice, we took the full 2D EEG data (only the S1 stimulus) and computed $\widehat{\mat{\Gamma}}^{\mathrm{TSIR}}_1, \widehat{\mat{\Gamma}}^{\mathrm{TSIR}}_2$, $\widehat{\mat{\Sigma}}^{\mathrm{TSIR}}_1, \widehat{\mat{\Sigma}}^{\mathrm{TSIR}}_2$ from which we computed the reductions $\widehat{\mat{\beta}}^{\mathrm{TSIR}}_1$ and $\widehat{\mat{\beta}}^{\mathrm{TSIR}}_2$. For comparison, we used GMLM to estimate the reduction matrices $\widehat{\mat{\beta}}^{\mathrm{GMLM}}_1$, $\widehat{\mat{\beta}}^{\mathrm{GMLM}}_2$ and the scatter matrices $\widehat{\mat{\Omega}}^{\mathrm{GMLM}}_1$, $\widehat{\mat{\Omega}}^{\mathrm{GMLM}}_2$. We also computed the GMLM analog of the standardized reductions $\widehat{\mat{\Gamma}}^{\mathrm{GMLM}}_1 = (\widehat{\mat{\Omega}}^{\mathrm{GMLM}}_1)^{-1}\widehat{\mat{\beta}}_1^{\mathrm{GMLM}}$ and $\widehat{\mat{\Gamma}}^{\mathrm{GMLM}}_2 = (\widehat{\mat{\Omega}}^{\mathrm{GMLM}}_2)^{-1}\widehat{\mat{\beta}}_2^{\mathrm{GMLM}}$. We next plotted in \cref{fig:TSIRvsGMLM} the estimated reductions $\widehat{\mat{\beta}}^{\mathrm{GMLM}}_1$, $\widehat{\mat{\beta}}^{\mathrm{GMLM}}_2$, $\widehat{\mat{\Gamma}}^{\mathrm{GMLM}}_1$, $\widehat{\mat{\Gamma}}^{\mathrm{GMLM}}_2$ in black and  $\widehat{\mat{\beta}}^{\mathrm{TSIR}}_1$, $\widehat{\mat{\beta}}^{\mathrm{TSIR}}_2$, $\widehat{\mat{\Gamma}}^{\mathrm{TSIR}}_1$, $\widehat{\mat{\Gamma}}^{\mathrm{TSIR}}_2$ in red (with appropriate scaling). Since the estimates of the reductions $\widehat{\mat{\beta}}_j$'s are completely different between GMLM and TSIR but the estimates of the standardized reductions $\mat{\Gamma}_j$'s agree to a high degree, we conclude that the loss of estimation accuracy happens in the final calculations that %That is in computing the estimates of the reductions $\mat{\beta}_j = \mat{\Sigma}_j^{-1}\mat{\Gamma}$, which 
involve the inversion of the severely ill-conditioned mode-wise estimated covariances $\widehat{\mat{\Sigma}}^{\mathrm{TSIR}}_1$ and $\widehat{\mat{\Sigma}}^{\mathrm{TSIR}}_2$.

To check whether our conjecture holds, we introduced a regularization term similar to the regularization employed in GMLM to the mode-wise sample covariances $\mat{\Sigma}_j$ in TSIR. As can be seen in \cref{fig:TSIRregvsGMLM}, this produced roughly the same reductions $\widehat{\mat{\beta}}_j$ as well as the $\widehat{\mat{\Gamma}}_j$.

\begin{figure}
\caption{TSIR (reg) in red vs. GMLM in black}
\label{fig:TSIRregvsGMLM}
    \includegraphics[width = \textwidth]{plots/tsir_regularized.pdf}
\end{figure}

To further study the relationship between GMLM and TSIR, we first  consider a setting in which GMLM and TSIR are theoretically equivalent. We assume that $Y$ is categorical, $\ten{X}\mid Y = y$ is multi-linear normal and the reduction $\ten{R}(\ten{X})$ is one-dimensional. Additionally, $\mat{\Omega}_j$ is assumed to be positive definite but \emph{not} further constrained. Under these assumptions, GMLM and TSIR are equivalent since %Even though GMLM minimizes the log-likelihood and TSIR minimizes a squared error loss, 
the theoretical mode-wise reduction matrices minimizing the log-likelihood (GMLM's objective) and the squared error loss (TSIR's objective) are theoretically identical.

In practice though, even in this simplest of settings, both procedures do not agree fully in their estimates. This is based on multiple different factors.
\begin{enumerate}
    \item Both procedures use different parameterizations. The reduction matrices $\mat{\beta}_j$ are parameters in GMLM in contrast to TSIR. As a result, TSIR first estimates its parameters $\mat{\Gamma}_j$ and mode-wise covariances $\mat{\Sigma}_j$ (corresponding to the marginal mode-wise covariances $\cov(\ten{X}_{(j)}$) and then estimates $\mat{\beta}_j$ with  $\widehat{\mat{\beta}}_j =\widehat{\mat{\Sigma}}_j^{-1}\widehat{\mat{\Gamma}}_j$. This ``two-step'' approach induces accumulation of estimation errors. Moreover, if any $\widehat{\mat{\Sigma}}_j$ is ill-conditioned, the resulting $\widehat{\mat{\beta}}_j$ is very unstable. In contrast, GMLM incorporates the reductions $\mat{\beta}_j$ directly into the model and estimates both $\mat{\beta}_j$ and $\mat{\Omega}_j$ at the same time, so that the estimation error in $\mat{\Omega}_j$ does \emph{not} affect the reductions. %But, in contrast to TSIR, GMLM does estimate both $\mat{\beta}_j$ and $\mat{\Omega}_j$ at the same time, linking the estimation accuracy of both in a different way. 
    This is due to GMLM's mode-wise scatter matrices $\mat{\Omega}_j$ are estimated from the inverse regression residuals (corresponding to $(\E\cov(\ten{X}_{(j)}\mid Y))^{-1}$).
    
    \item The original version of TSIR \cite[]{DingCook2015} does \emph{not} incorporate any regularization. We introduced regularization to TSIR leading to the TSIR (reg) procedure by simply adding $0.2\lambda_1(\mat{\Sigma}_j)\mat{I}_{p_j}$ to $\mat{\Sigma}_j$ if and only if the condition number of $\mat{\Sigma}_j$ exceeds an arbitrary threshold of $25$, similar to the regularization used in GMLM. Without this regularization, TSIR can fail dramatically.
    %in the final computation $\mat{\beta}_j = \mat{\Sigma}_j^{-1}\mat{\Gamma}_j$, which is the reason for the drop in AUC in \cref{tab:eeg} for TSIR in contrast to TSIR (reg). 
    In comparison, GMLM behaves more stable in regard to estimating the reductions $\mat{\beta}_j$ even without regularization. At the same time, the need for regularization is more pronounced in GMLM. This means that even in well behaved settings, the estimation algorithm for GMLM does rely on its regularization to guard against divergence in the iterative estimation procedure. In summary, without regularization, TSIR may encounter complete breakdown without regularization while GMLM has worse average performance (not directly reported since the regularization in GMLM is the default for the multi-linear normal algorithm, applied only in a subset of iterations).
\end{enumerate}

If the response $Y$ is a univariate continuous variable, then GMLM and TSIR are no longer equivalent (from a theoretical point of view.)

\begin{enumerate}\setcounter{enumi}{2}
    \item TSIR performs slicing of the continuous response $y$ which misses fine-grained information. GMLM does not have this restriction, being able to directly work with the continuous response. This is the main reason for the slight performance gain of GMLM over TSIR in the simulations of \cref{sec:sim-tensor-normal} (due to the small dimensionality of the simulations, regularization for TSIR is not needed while GMLM applies regularization during its iterative estimation procedure for algorithmic stability only. Meaning, the final GMLM estimates are also almost never regularized).
    \item Another major difference between GMLM and TSIR is the tensor-valued function $\ten{F}_y$ of the response, which makes GMLM much more general than TSIR. If $\ten{F}_y$ is the fraction of observations in a slice, TSIR is in effect a special case of GMLM.  %During the modeling process of GMLM we introduced .% $y$. 
    %This was assumed as a \emph{known} function of $y$, which in practice is subject to modeling. This is another major difference between GMLM and TSIR, manifesting in two parts.
    
    For general $\ten{F}_y\in\mathbb{R}^{q_1\times\ldots q_r}$, its dimension, $q = \prod_{j = 1}^r q_j$,   inherently constrains the dimension of the reduction in \cref{thm:sdr}.  TSIR also requires to provide a per-mode reduction dimension $\tilde{q}_j$ which leads to a final reduction of dimensionality $\tilde{q} = \prod_{j = 1}^r \tilde{q}_j$, but without a link to an explicitly provided function. This makes the specification of the reduction dimension simpler, while being equality constraint from a number theoretic divisibility point of view. 
    
\end{enumerate}

\noindent So far, we required no additional constraints on the model parameters.

\begin{enumerate}\setcounter{enumi}{4}
    \item Given additional information about the regression at hand, one might want to incorporate that knowledge into the method. For example, for longitudinal data, the time dimension may be modeled as an AR process. GMLM allows (under the conditions outlined in \cref{sec:manifolds}) such modeling by describing the corresponding parameter manifolds of the covariances or the reductions themselves, a functionality \emph{not} available in TSIR.
\end{enumerate}


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Parameter estimation for the Multi-Linear Ising model}\label{sec:ising-param-estimation}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Initial Values}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
The first step is to get reasonable starting values. Experiments showed that a good starting value of $\mat{\beta}_k$ is to use the multi-linear normal estimates from \cref{sec:tensor-normal-estimation} for $k = 1, \ldots, r$, considering $\ten{X}_i$ as continuous. For initial values of $\mat{\Omega}_k$, a different approach is required. Setting everything to the uninformative initial value results in  $\mat{\Omega}_k = \mat{0}$ as this corresponds to the conditional log odds to be $1:1$ for every component and pairwise interaction. This is not possible, since $\mat{0}$ is a stationary point of the log-likelihood, as can be directly observed by considering the partial gradients of the log-likelihood in \cref{thm:grad}. Instead, we use a crude heuristic that threads every mode separately and ignores any relation to the covariates. It is computationally cheap and better than any of the alternatives we considered. For every $k = 1, \ldots, r$, let the $k$th mode second moment estimate be
\begin{equation}\label{eq:ising-mode-moments}
    \hat{\mat{M}}_{2(k)} = \frac{p_k}{n p}\sum_{i = 1}^n (\ten{X}_i)_{(k)}\t{(\ten{X}_i)_{(k)}}
\end{equation}
which contains the $k$th mode first moment estimate in its diagonal $\hat{\mat{M}}_{1(k)} = \diag\hat{\mat{M}}_{2(k)}$.

For the purpose of an initial value estimate, we threat each of the columns of the matricized observation $(\ten{X}_i)_{(k)}$ as an i.i.d. observation of a $p_k$ dimensional random variable $Z_k$. Through this reasoning we get with $n$ i.i.d. observations $\ten{X}_i$ a total of $n \prod_{j\neq k}p_j$ realizations of the random variable $Z_k$ for each of the modes $k = 1, \ldots, r$.

Continuing this reasoning, the elements of $(\hat{\mat{M}}_{1(k)})_{j}$ are the estimates of the marginal probability $P((Z_k)_j = 1)$ of the $j$th element of $Z_k$ being $1$. Similarly, for $l \neq j$, the entry $(\hat{\mat{M}}_{2(k)})_{j l}$ estimates the marginal probability of two-way interactions, $P((Z_k)_j = 1, (Z_k)_l = 1)$. Now, we set the diagonal elements of $\mat{\Omega}_k$ to zero. For the off diagonal elements of $\mat{\Omega}_k$, we equate the conditional probabilities $P((Z_k)_j = 1 \mid (Z_k)_{-j} = \mat{0})$ and $P((Z_k)_j = 1, (Z_k)_l = 1\mid (Z_k)_{-j, -l} = \mat{0})$ with the marginal probability estimates $(\hat{\mat{M}}_{1(k)})_{j}$ and $(\hat{\mat{M}}_{2(k)})_{j l}$, respectively. Applying \eqref{eq:ising-two-way-log-odds}  gives the initial component-wise estimates $\hat{\mat{\Omega}}_k^{(0)}$, 
\begin{equation}\label{eq:ising-init-Omegas}
    (\hat{\mat{\Omega}}_k^{(0)})_{j j} = 0,
    \qquad
    (\hat{\mat{\Omega}}_k^{(0)})_{j l} = \log\frac{1 - (\hat{\mat{M}}_{1(k)})_{j}(\hat{\mat{M}}_{1(k)})_{l}}{(\hat{\mat{M}}_{1(k)})_{j}(\hat{\mat{M}}_{1(k)})_{l}}\frac{(\hat{\mat{M}}_{2(k)})_{j l}}{1 - (\hat{\mat{M}}_{2(k)})_{j l}}, \, j \neq l.
\end{equation}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Gradient Optimization}
Given initial values, the gradients derived in  \cref{thm:grad} can be evaluated for the Ising model. The first step therefore is to determine the values of the inverse link components $\ten{g}_1(\mat{\gamma}_y) = \E[\ten{X} \mid Y = y]$ and $\ten{G}_2(\mat{\gamma}_y) = \ten{g}_2(\mat{\gamma}_y) = \E[\ten{X}\circ\ten{X} \mid Y = y]$. An immediate simplification is that the first moment is  part of the second moment. Its values are determined via $\vec(\E[\ten{X} \mid Y = y]) = \diag(\E[\ten{X}\circ\ten{X} \mid Y = y]_{(1, \ldots, r)})$; i.e.,  only the second moment needs to be computed, or estimated (see \cref{sec:ising-bigger-dim}) in the case of slightly bigger $p$. For the Ising model, the conditional second moment with parameters $\mat{\gamma}_y$ is given by the matricized relation
\begin{equation}\label{eq:ising-m2}
    \ten{g}_2(\ten{\gamma}_y)_{(1, \ldots, r)} = \E\left[(\vec{\ten{X}})\t{(\vec{\ten{X}})}\mid Y = y\right] = p_0(\mat{\gamma}_y)\sum_{\mat{x}\in\{0, 1\}^{p}}\mat{x}\t{\mat{x}}\exp(\t{\vech(\mat{x}\t{\mat{x}})}\mat{\gamma}_y).
\end{equation}
The natural parameter $\mat{\gamma}_y$ is evaluated via \eqref{eq:ising-natural-params} enabling us to compute the partial gradients of the log-likelihood $l_n$ \eqref{eq:log-likelihood} for the Ising model by \cref{thm:grad} for the GMLM parameters $\mat{\beta}_k$ and $\mat{\Omega}_k$, $k = 1, \ldots, r$, at the current iterate $\mat{\theta}^{(I)} = (\mat{\beta}_1^{(I)}, \ldots, \mat{\beta}_r^{(I)}, \mat{\Omega}_1^{(I)}, \ldots, \mat{\Omega}_r^{(I)})$. Using classic gradient ascent for maximizing the log-likelihood, we have to specify a learning rate $\lambda\in\mathbb{R}_{+}$, usually a value close to $10^{-3}$. The update rule is
\begin{displaymath}
    \mat{\theta}^{(I + 1)} = \mat{\theta}^{(I)} + \lambda\nabla_{\mat{\theta}} l_n(\mat{\theta})\bigr|_{\mat{\theta} = \mat{\theta}^{(I)}},
\end{displaymath}
which is iterated till convergence. In practice, iteration is performed until either a maximum number of iterations is exhausted and/or some break condition is satisfied. A proper choice of the learning rate is needed as a large learning rate $\lambda$ may cause instability, while a very low learning rate requires an enormous amount of iterations. Generically, there are two approaches to avoid the need to determine a proper learning rate. First, \emph{line search methods} determine an appropriate step size for every iteration. This works well if the evaluation of the object function (the log-likelihood) is cheap. This is not the case in our setting, see \cref{sec:ising-bigger-dim}. The second approach is an \emph{adaptive learning rate}, where one tracks specific statistics while optimizing and dynamically adapting the learning rate via well-tested heuristics using the gathered knowledge from past iterations. We opted to use an adaptive learning rate approach, which not only removes the need to determine an appropriate learning rate but also accelerates learning.

Our method of choice is \emph{root mean squared propagation} (RMSprop) \cite[]{Hinton2012}. This is a well-known method in machine learning for training neural networks. It is a variation of gradient descent with a per scalar parameter adaptive learning rate. It tracks a moving average of the element-wise squared gradient $\mat{g}_2^{(I)}$, which is then used to scale (element-wise) the gradient in the update rule (see  \cite{Hinton2012} and \cite{GoodfellowEtAl2016} among others). The update rule using RMSprop for maximization\footnote{Instead of the more common minimization, therefore $+$ in the update of $\mat{\theta}$.} is
\begin{align*}
    \mat{g}_2^{(I + 1)} &= \nu \mat{g}_2^{(I)} + (1 - \nu)\nabla l_n(\mat{\theta}^{(I)})\odot\nabla l_n(\mat{\theta}^{(I)}), \\
    \mat{\theta}^{(I + 1)} &= \mat{\theta}^{(I)} + \frac{\lambda}{\sqrt{\mat{g}_2^{(I + 1)}} + \epsilon}\odot\nabla l_n(\mat{\theta}^{(I)}).
\end{align*}
The parameters $\nu = 0.9$, $\lambda = 10^{-3}$ and $\epsilon\approx 1.49\cdot 10^{-8}$ are fixed. The initial value of $\mat{g}_2^{(0)} = \mat{0}$, where the symbol $\odot$ denotes the Hadamard product, or element-wise multiplication. The division and square root operations are performed element-wise as well. According to our experiments, RMSprop requires iterations in the range of $50$ till $1000$ till convergence while gradient ascent with a learning rate of $10^{-3}$ is in the range of $1000$ till $10000$.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Small Data Sets}\label{sec:ising-small-data-sets}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
In the case of a finite number of observations, specifically in data sets with a small number of observations $n$, the situation where one component is always either zero or one can occur. It is also possible to observe two exclusive components. In practice, this situation of a ``degenerate'' data set should be protected against. Working with parameters on a log scale, gives estimates of $\pm\infty$, which is outside the parameter space and breaks our optimization algorithm.

The first situation where this needs to be addressed is in \eqref{eq:ising-init-Omegas}, where we set initial estimates for $\mat{\Omega}_k$. To avoid division by zero as well as evaluating the log of zero, we adapt \eqref{eq:ising-mode-moments}, the mode-wise moment estimates $\hat{\mat{M}}_{2(k)}$. A simple method is to replace the ``degenerate'' components, that are entries with value $0$ or $1$, with the smallest positive estimate of exactly one occurrence $p_k / n p$, or all but one occurrence $1 - p_k / n p$, respectively.

The same problem is present in gradient optimization. Therefore, before starting the optimization, we detect degenerate combinations. We compute upper and lower bounds for the ``degenerate'' element in the Kronecker product $\hat{\mat{\Omega}} = \bigkron_{k = r}^{1}\hat{\mat{\Omega}}_k$. After every gradient update, we check if any of the ``degenerate'' elements fall outside of the bounds. In that case, we adjust all the elements of the Kronecker component estimates $\hat{\mat{\Omega}}_k$, corresponding to the ``degenerate'' element of their Kronecker product, to fall inside the precomputed bounds. While doing so, we try to alter every component as little as possible to ensure that the non-degenerate elements in $\hat{\mat{\Omega}}$, affected by this change due to its Kronecker structure, are altered as little as possible. The exact details are technically cumbersome while providing little insight.


%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\subsection{Slightly Bigger Dimensions}\label{sec:ising-bigger-dim}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
A big challenge for the Ising model is its high computational complexity as it involves summing over all binary vectors of length $p = \prod_{k = 1}^{r}p_k$ in the partition function \eqref{eq:ising-partition-function}. Exact computation of the partition function requires summing all $2^p$ binary vectors. For small dimensions, say $p\approx 10$, this is easily computed. Increasing the dimension beyond $20$ becomes extremely expensive and  %absolutely
impossible for a dimension bigger than $30$. Trying to avoid the evaluation of the log-likelihood and only computing its partial gradients via \cref{thm:grad} does not resolve the issue. The gradients require the inverse link, that is the second moment \eqref{eq:ising-m2}, which still involves summing $2^p$ terms if  the scaling factor $p_0$ is dropped. Basically, with our model, this means that the optimization of the Ising model using exactly computed gradients is impossible for moderately sized problems.

When $p=\prod_{i=1}^r p_i > 20$, we use a Monte-Carlo method to estimate the second moment \eqref{eq:ising-m2}, required to compute the partial gradients of the log-likelihood. Specifically, we use a Gibbs-Sampler to sample from the conditional distribution and approximate the second moment in an importance sampling framework. This can be implemented quite efficiently and the estimation accuracy for the second moment is evaluated experimentally. %which seems to be very reliable. 
Simultaneously, we use the same approach to estimate the partition function. This, though, is inaccurate and may only be used to get a rough idea of the log-likelihood. Regardless, for our method, we only need the gradient for optimization where appropriate break conditions, not based on the likelihood, lead to a working method for MLE estimation.

\begin{figure}[!hpt]
    \centering 
    \includegraphics[width=0.5\textwidth]{../plots/sim-ising-perft-m2.pdf} 
    \caption{\label{fig:ising-m2-perft}Performance test for computing/estimating the second moment of the Ising model of dimension $p$ using either the exact method or a Monte-Carlo (MC) simulation.}
\end{figure}



%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
\section{Chess}\label{sec:chess}
%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
We provide a \emph{proof of concept} analysis of chess positions based on the Ising GMLM model where we build a mixture of matrix Ising models. Our modeling is based on analyzed games to build a sufficient reduction for winning positions.

The data set is provided by the \emph{lichess.org open database}.\footnote{\emph{lichess.org open database}. visited on December 8, 2023. \url{https://database.lichess.org}} We randomly selected the November of 2023 data that consist of more than $92$ million games. We removed all games without position evaluations. The evaluations, also denoted as scores, are from Stockfish,\footnote{The Stockfish developers [see \href{https://github.com/official-stockfish/Stockfish/blob/master/AUTHORS}{AUTHORS} file] (since 2008). \textit{Stockfish}. Stockfish is a free and strong UCI chess engine. \url{https://stockfishchess.org}} a free and strong chess engine. The scores take the role of the response $Y$ and correspond to a winning probability from the white pieces' point of view. Positive scores are good for white and negative scores indicate an advantage for black pieces. We ignore all highly unbalanced positions, which we set to be positions with absolute score above $5$. We also remove all positions with a mate score (one side can force checkmate). Furthermore, we only consider positions after $10$ half-moves to avoid oversampling the beginning of the most common openings including the start position which is in every game. Finally, we only consider positions with white to move. This leads to a final data set of roughly $64$ million positions, including duplicates. We did not filter according to the players strength nor the time control of the game. This means that our data set contains games from very weak to very strong players as well as very fast games, like Bullet, to classic chess.

A chess position is encoded as a set of $12$ binary matrices $\ten{X}_{\mathrm{piece}}$ of dimensions $8\times 8$. Every binary matrix encodes the positioning of a particular piece by containing a $1$ if the piece is present at the corresponding board position. The $12$ pieces derive from the $6$ types of pieces, namely pawns (\pawn), knights (\knight), bishops (\bishop), rooks (\rook), queens (\queen), and kings (\king) of two colors, black and white. See \cref{fig:fen2tensor} for a visualization.

\begin{figure}[!hpt]
    \centering
    %\includegraphics[width = \textwidth]{../images/fen2tensor.pdf}
     \includegraphics[scale=0.8]{../images/fen2tensor.pdf}\caption{\label{fig:fen2tensor}The chess start position and its 3D binary tensor representation, empty entries are $0$.}
\end{figure}

We assume that $\ten{X}_{\mathrm{piece}}\mid Y = y$ follows an Ising GMLM model as in \cref{sec:ising_estimation} with different conditional piece predictors being independent. %The independence assumption is for the sake of simplicity even though this is not the case in the underlying true distribution. 
The simplifying assumption of independence results in a mixture model with  log-likelihood
\begin{displaymath}
    l_n(\mat{\theta}) = \frac{1}{12}\sum_{\mathrm{piece}}l_n(\mat{\theta}_{\mathrm{piece}})
\end{displaymath}
where $l_n(\mat{\theta}_{\mathrm{piece}})$ is the Ising GMLM log-likelihood as in \cref{sec:ising_estimation} for $\ten{X}_{\mathrm{piece}}\mid Y = y$. For every component the same relation to the scores $y$ is modeled via a $2\times 2$ dimensional matrix-valued function $\ten{F}_y$ consisting of the monomials $1, y, y^2$, specifically $(\ten{F}_y)_{i j} = y^{i + j - 2}$.

Due to the volume of the data  (millions of observations), it is  computationally infeasible to compute the gradients on the entire data set. %Simply using a computationally manageable subset is not an option. 
The dimension of the  binary data is $12$ times a $8\times 8 =768$ for every observation. %The main issue is that a manageable subset, say one million observations, still leads to a degenerate data set. In our simplified mixture model, the pawns are a specific issue as there are multiple millions of different combinations of the $8$ pawns per color on the $6\times 8$ sub-grid where the pawns can be positioned. This alone does not allow us to take a reasonable sized subset for estimation. 
The solution is to switch from a classic gradient-based optimization to a stochastic version; that is, every gradient update uses a new random subset of the entire data set. We draw independent random samples from the data consisting of $64$ million positions. The independence of samples is derived from the independence of games, and every sample is drawn from a different game.

\subsection{Validation}
Given the non-linear nature of the reduction, due to the quadratic matrix-valued function $\ten{F}_y$ of the score $y$, we use a \emph{generalized additive model}\footnote{using the function \texttt{gam()} from the \texttt{R} package \texttt{mgcv}.} (GAM) to predict position scores from reduced positions. The reduced positions are $48$ dimensional continuous values by combining the $12$ mixture components from the $2\times 2$ matrix-valued reductions per piece. The per-piece reduction is
\begin{displaymath}
    \ten{R}(\ten{X}_{\mathrm{piece}}) = \mat{\beta}_{1,\mathrm{piece}}(\ten{X}_{\mathrm{piece}} - \E\ten{X}_{\mathrm{piece}})\t{\mat{\beta}_{2, \mathrm{piece}}}
\end{displaymath}
which gives the complete $48$ dimensional vectorized reduction by stacking the piece-wise reductions
\begin{displaymath}
    \vec{\ten{R}(\ten{X}})
        = (\vec{\ten{R}(\ten{X}_{\text{white pawn}})}, \ldots, \vec{\ten{R}(\ten{X}_{\text{black king}})})
        = \t{\mat{B}}\vec(\ten{X} - \E\ten{X}).
\end{displaymath}
The second line encodes all the piece-wise reductions in a block diagonal full reduction matrix $\mat{B}$ of dimension $768\times 48$ which is applied to the vectorized 3D tensor $\ten{X}$ combining all the piece components $\ten{X}_{\mathrm{piece}}$ into a single tensor of dimension $8\times 8\times 12$. This is a reduction to $6.25\%$ of the original dimension. The $R^2$ statistic of the GAM fitted on $10^5$ new reduced samples is $R^2_{\mathrm{gam}}\approx 46\%$. A linear model on the reduced data achieves $R^2_{\mathrm{lm}}\approx 26\%$ which clearly shows the non-linear relation. On the other hand, the static evaluation of the \emph{Schach H\"ornchen}\footnote{Main author's chess engine.} engine, given the full position (\emph{not} reduced), achieves an $R^2_{\mathrm{hce}}\approx 52\%$. The $42\%$ are reasonably well compared to $51\%$ of the engine static evaluation which gets the original position and uses chess specific expert knowledge. Features the static evaluation includes, which are expected to be learned by the GMLM mixture model, are the  \emph{material} (piece values) and \emph{piece square tables} (PSQT, preferred piece type positions). In addition, the static evaluation includes chess specific features like \emph{king safety}, \emph{pawn structure}, or \emph{rooks on open files}. This lets us conclude that the reduction captures most of the relevant features possible, given the oversimplified modeling we performed.

\subsection{Interpretation}
For a compact interpretation of the estimated reduction we construct PSQTs. To do so we use the linear model from the validation section. Then, we rewrite the combined linear reduction and linear model in terms of PSQTs. Due to the nature of our analysis, it does not provide the usual PSQT interpreted as the square where a piece is most powerful, but instead that the presence of a piece on a particular square is indicative of a winning position. Those two are different in the sense that the first indicates the potential of a piece on a particular square while the second also incorporates the cases where this potential was already used, resulting in a winning position (often due to an oversight of the opponent).

Let $\mat{B}$ be the $768\times 48$ full vectorized linear reduction. This is the block diagonal matrix with the $64\times 4$ dimensional per piece reductions $\mat{B}_{\mathrm{piece}} = \mat{\beta}^{\mathrm{piece}}_2\otimes\mat{\beta}^{\mathrm{piece}}_1$. Then, the linear model with coefficients $\mat{b}$ and intercept $a$ on the reduced data is given by
\begin{equation}\label{eq:chess-lm}
    y = a + \t{\mat{b}}\t{\mat{B}}\vec(\ten{X} - \E\ten{X}) + \epsilon
\end{equation}
with an unknown mean zero error term $\epsilon$ and treating the binary tensor $\ten{X}$ as continuous. Decomposing the linear model coefficients into blocks of $4$ gives per piece coefficients $\mat{b}_{\mathrm{piece}}$ which combine with the diagonal blocks $\mat{B}_{\mathrm{piece}}$ of $\mat{B}$ only. Rewriting \eqref{eq:chess-lm} gives
\begin{align*}
    y &= a + \sum_{\mathrm{piece}}\t{(\mat{B}_{\mathrm{piece}}\mat{b}_{\mathrm{piece}})}\vec(\ten{X}_{\mathrm{piece}} - \E\ten{X}_{\mathrm{piece}}) + \epsilon \\
    &= \tilde{a} + \sum_{\mathrm{piece}}\langle
        \mat{B}_{\mathrm{piece}}\mat{b}_{\mathrm{piece}},
        \vec(\ten{X}_{\mathrm{piece}})
    \rangle + \epsilon
\end{align*}
with a new intercept term $\tilde{a}$, which is of no interest to us. Finally, we enforce color symmetry, using known mechanisms from chess engines. Specifically, mirroring the position changes the sign of the score $y$. Here, mirroring reverses the rank (row) order, this is the image in a mirror behind a chess board. Let every $\mat{C}_{\mathrm{piece}}$ be a $8\times 8$ matrix with elements $(\mat{C}_{\mathrm{piece}})_{i j} = (\mat{B}_{\mathrm{piece}}\mat{b}_{\mathrm{piece}})_{i + 8 (j - 1)}$. And denote with $\mat{M}(\mat{A})$ the matrix mirror operation which reverses the row order of a matrix.
The new notation allows to enforce the symmetry, resulting in an approximate linear relationship.
\begin{align*}
    y &= \tilde{a} + \sum_{\mathrm{piece}}\langle
        \mat{C}_{\mathrm{piece}},
        \ten{X}_{\mathrm{piece}}
        \rangle + \epsilon \\
    &\approx \tilde{a} + \sum_{\text{piece type}}\frac{1}{2}\langle
        \mat{C}_{\text{white piece}} - \mat{M}(\mat{C}_{\text{black piece}}),
        \ten{X}_{\text{white piece}} - \mat{M}(\ten{X}_{\text{white piece}})
    \rangle + \epsilon
\end{align*}
If $\mat{C}_{\text{white piece}} = -\mat{M}(\mat{C}_{\text{black piece}})$ for every piece type, then we have equality ($6$ types, \emph{not} distinguishing between color.) In our case, this is valid given that the estimates $\hat{\mat{C}}_{\mathrm{piece}}$ fulfill this property with a small error. The $6$ matrices $(\mat{C}_{\text{white piece}} - \mat{M}(\mat{C}_{\text{black piece}})) / 2$ are called \emph{piece square tables} (PSQT) which are visualized in \cref{fig:psqt}. The interpretation of those tables is straightforward. A high positive value (blue) means that it is usually good to have a piece of the corresponding type on that square while a high negative value (red) means the opposite. It needs to be considered that the PSQTs are for quiet positions only, which means all pieces are save in the sense that there is no legal capturing moves nor is the king in check.

\begin{figure}[t]
    \centering
    %\includegraphics[width = \textwidth]{../plots/psqt.pdf}
    \includegraphics[scale=0.8]{../plots/psqt.pdf}
    \caption{\label{fig:psqt}Extracted PSQTs (piece square tables) from the chess example GMLM reduction.}
\end{figure}

The first visual effect in \cref{fig:psqt} is the dark blue PSQT of the Queen followed by a not-so-dark Rook PSQT. This indicates that the Queen, followed by the Rook, are the most valuable pieces (after the king which is the most valuable, which also implies that assigning value to the king makes no sense). The next two are the Knight and Bishop which have higher value than the Pawns, ignoring the $6$th and $7$th rank as this makes the pawns potential queens. This is the classic piece value order known in chess. 

Next, going over the PSQTs one by one, a few words about the preferred positions for every piece type. The pawn positions are specifically good on the $6$th and especially on the $7$th rank as this threatens a promotion to a Queen (or Knight, Bishop, Rook). The Knight PSQT is a bit surprising, the most likely explanation for the knight being good in the enemy territory is that it got there by capturing an enemy piece for free. A common occurrence in low-rated games is a big chunk of the training data, ranging over all levels. The Bishops seem to have no specific preferred placement, only a slightly higher overall value than pawns, excluding pawns imminent for a promotion. Continuing with the rooks, we see that the rook is a good attacking piece, indicated by a save rook infiltration\footnote{Rook infiltration is a strategic concept in chess that involves skillfully maneuvering your rook to penetrate deep into your opponent's territory.}. It is well known chess knowledge that the Rook is most powerful on the 7th rank, but this is different from our analysis as this is related to the potential of a Rook in contrast to the position of a Rook in a winning position over all states of the game. The Queen is powerful almost everywhere, only the outer back rank squares (lower left and right) tend to reduce her value. This is rooted in the queen's presence there being a sign for being chased by enemy pieces. Leading to a lot of squares being controlled by the enemy hindering one own movement. Finally, given the goal of the game is to checkmate the king, a safe position for the king is very valuable. This is seen by the back rank (rank $1$) being the only non-penalized squares. Furthermore, the safest squares are the castling\footnote{Castling is a maneuver that combines king safety with rook activation.} target squares ($g1$ and $c1$) as well as the $b1$ square. Shifting the king over to $b1$ is quite common to protect the $a2$ pawn so that the entire pawn shield in front of the king is protected.

The results of our analysis above agree with the configuration of the chess board most associated with observed chess game outcomes. This analysis also aligns with the understanding of human chess players of an average configuration at any moment during the game when considering the different interpretation of being in a winning position in contrast to the power of a piece on a specific square. Moreover, our analysis was performed over a wide range of player skill levels while disregarding if a position is in the middle or end game.

\end{appendix}

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%\begin{acks}[Acknowledgments]
%The authors would like to thank the anonymous referees, an Associate Editor and the Editor for their constructive comments that improved the quality of this paper.
%\end{acks}

%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%%
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\begin{funding}
Both authors were partially supported by the Vienna Science and Technology Fund (WWTF) [10.47379/ICT19018] and the Austrian Science Fund (FWF) research
project P 30690-N35.
\end{funding}

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