# Marginalizable Density Models

```Marginalizable Density Models

Dar Gilboa                                    Ari Pakman
Harvard University                            Columbia University
dar_gilboa@fas.harvard.edu                       ari@stat.columbia.edu

Thibault Vatter
arXiv:2106.04741v1 [stat.ML] 8 Jun 2021

Columbia University
thibault.vatter@columbia.edu

Abstract
Probability density models based on deep networks have achieved remarkable
success in modeling complex high-dimensional datasets. However, unlike kernel
density estimators, modern neural models do not yield marginals or conditionals in
closed form, as these quantities require the evaluation of seldom tractable integrals.
In this work, we present the marginalizable density model approximator (MDMA),
a novel deep network architecture which provides closed form expressions for the
probabilities, marginals and conditionals of any subset of the variables. The MDMA
learns deep scalar representations for each individual variable and combines them
via learned hierarchical tensor decompositions into a tractable yet expressive CDF,
from which marginals and conditional densities are easily obtained. We illustrate
the advantage of exact marginalizability in several tasks that are out of reach of
previous deep network-based density estimation models, such as estimating mutual
information between arbitrary subsets of variables, inferring causality by testing
for conditional independence, and inference with missing data without the need for
data imputation, outperforming state-of-the-art models on these tasks. The model
also allows for parallelized sampling with only a logarithmic dependence of the
time complexity on the number of variables.

1   Introduction
Estimating the joint probability density of a set of random variables is a fundamental task in statistics
and machine learning that has witnessed much progress in recent years. While traditional estimators
such as histograms [1, 2] and kernel-based methods [3, 4] have appealing theoretical properties and
typically perform well in low dimensions, they become computationally impractical above 5-10
dimensions. Conversely, recent density models based on neural networks [5–10] scale efficiently
with the number of random variables, but lack a crucial feature available to traditional methods: the
ability to compute probabilities, marginalize over subsets of variables, and evaluate conditionals.
These tasks require integrals of the estimated density, which are intractable for modern neural density
models. Thus, while such operations are central to many applications (e.g., inference with missing
data, testing for conditional (in)dependence, or performing do-calculus ), approaches based on
neural networks require estimating separate models whenever marginals or conditionals are needed.
Alternatively, one could model the cumulative distribution function (CDF), making computing
probabilities and marginalization straightforward. But evaluating the density requires taking d
derivatives of the CDF, which incurs an exponential cost in d for a generic computational graph. This
In this work, we present the marginalizable density model approximator (MDMA), a novel deep
network architecture preserving most of the expressive power of neural models for density estimation,

Preprint. Under review.```
```while providing closed form expressions for the probabilities, marginals and conditionals of any
subset of the variables. In a nutshell, the MDMA learns many deep scalar representations for each
individual variable and combines them using hierarchical tensor decompositions [13, 14] into a
tractable multivariate CDF that can be fitted using stochastic gradient descent. Additionally, sampling
from MDMA can be parallelized along the input dimension, resulting in a very low space complexity
and a time complexity that scales only logarithmically with the number of variables in the problem
(as opposed to linearly in naive autoregressive sampling, see below in Section 2).
As could be expected, the architectural choices that allow for easy marginalization take a minor toll in
terms of performance. Indeed, while competitive, our model admittedly does not beat state-of-the-art
models in out-of-sample log-likelihood of high-dimensional datasets. On the other hand, it does
beat those same models in a task for which the latter are ill-prepared: learning densities from data
containing missing values, a common setting in some application areas such as genomics [15, 16].
While our model is able to deal optimally with missing values by evaluating, for every data point,
the marginal likelihood over its non-missing values, other models must resort to data imputation.
Consequently, we significantly outperform state-of-the-art neural density estimators trained using
a number of common data imputation strategies. We also show MDMA can be used to test for
conditional independence, which is useful for discovering the causal structure in graphs, a task on
which it outperforms existing methods, and that it enables estimation of mutual information between
arbitrary subsets of variables after fitting a single model. Additionally, we prove that the model class
is a universal approximator over the space of absolutely continuous multivariate distributions.
The structure of this paper is as follows. In Section 2 we review related works. In Section 3 we
present our new model and its theoretical properties. We present our experimental results in Section 4,
and conclude in Section 5.

2     Related work
Modern approaches to non-parametric density estimation, based on normalizing flows [5–10] (see [17,
18] for recent reviews), model expressive yet invertible functions that transforms a simple (usually
uniform or Gaussian) density to the target density. Nonetheless, as previously mentioned, such
architectures lead to intractable derived quantities such as probabilities, marginals and/or conditional
distributions.
Moreover, many normalizing flow models rely on an autoregressive construction, which makes the
cost of generating samples scale linearly with the dimension. This can be circumvented using inverse
autoregressive flows , but in this dual case a linear cost is incurred instead in density evaluations
and hence in training. Another solution is training a feed-forward network using the outputs of a
trained autoregressive model . With MDMA, fast inference and sampling is achieved without
requiring this distillation procedure.
Tensor decompositions [13, 14], which are exploited in this work, have been used in various appli-
cations of signal processing, machine learning, computer vision, and more [21–25]. Recently, such
decompositions have been used to speed-up or reduce the number of parameters in existing deep
architectures [26–31]. In addition to their practical appeal, tensor methods have been widely studied
to understand the success of deep neural networks [32–36]

3     Marginalizable Density Models

3.1   Notations

In the following, we use a capital and lowercase Roman letter (e.g., F and f ) or Greek letters
along with dot above (e.g., ϕ and ϕ̇) to denote respectively absolutely continuous CDFs of arbitrary
dimensions and the corresponding densities. When dealing with multivariate distributions, the
marginal or conditional distribution over a subset of variables will be indicated by the argument
names (i.e., F (x1 |x2 , x3 ) = P [X1 ≤ x1 |X2 = x2 , X3 = x3 ]).
For a positive integer p ∈ N \ 0, let [p] = {1, . . . , p}. Denote the space of absolutely continuous
univariate and d-dimensional CDFs respectively by F1 and Fd . For any F ∈ F1 , the density
f : R → R+ is f (x) = ∂F (x)/∂x. Similarly, for any F ∈ Fd , the density f (x) : Rd → R+ is

2```
```f (x) = ∂ d F (x)/∂x1 · · · ∂xd , and Fj (x) = limz→∞ F (z, . . . , x, . . . , z) ∈ F1 for j ∈ [d] is the jth
marginal distribution.

3.2   The bivariate case

For the task of modeling joint distributions of two variables supported on R2 , consider a family of
univariate CDFs {ϕi,j }i∈[m], j∈ with ϕi,j ∈ F1 , i.e., the functions ϕi,j : R → [0, 1] satisfy
lim ϕi,j (x) = 0 ,             lim ϕi,j (x) = 1 ,               ϕ̇i,j (x) = ∂ϕ(x)/∂x ≥ 0 .
x→−∞                            x→∞

These functions are our basic building block, and we model them using a simple neural architecture
proposed in
P  and described in Section 3.6. If A is an m × m matrix of nonnegative elements
m
satisfying i,j=1 Ai,j = 1, we can combine it with the univariate CDFs to obtain
m
X
F (x1 , x2 ) =           Ai,j ϕi,1 (x1 )ϕj,2 (x2 ).                                 (1)
i,j=1

The coefficients Ai,j encode the dependencies between the two variables, and the normalization
ensures that F is a valid CDF, that is F ∈ F2 . Even though in each summand the interaction
is modeled by a single scalar parameter, such a model can be used to approximate well complex
interactions between x1 and x2 if m is sufficiently large, as we show in Section 3.6. The advantage
of this construction is that {ϕ̇i,j }i∈[m], j∈ , the family of densities corresponding to the univariate
m
X
f (x1 , x2 ) =           Ai,j ϕ̇i,1 (x1 )ϕ̇j,2 (x2 ).
i,j=1

It is similarly straightforward to obtain marginal and conditional quantities, e.g.:
m
P
m                                                            Ai,j ϕi,1 (x1 )ϕ̇j,2 (x2 )
X                                                  i,j=1
F (x1 ) =           Ai,j ϕi,1 (x1 ),      F (x1 |x2 ) =                  m
P                            ,
i,j=1                                                              Ai,j ϕ̇j,2 (x2 )
i,j=1

and the corresponding densities result from replacing ϕi,1 by ϕ̇i,1 . Deriving these simple expressions
relies on the fact that (1) combines the univariate CDFs linearly. Nonetheless, it is clear that, with
m → ∞ and for judiciously chosen univariate CDFs, such a model is a universal approximator of
both CDFs and sufficiently smooth densities.

3.3   The multivariate case

To generalize the bivariate case, consider a collection of univariate CDFs {ϕi,j }i∈[m], j∈[d] with
ϕi,j ∈ F1 for each i and j, and define the tensor-valued function Φ : Rd × [m]d → [0, 1] by
Qd
Φ(x)i1 ,...,id = j=1 ϕij ,j (xj ) for x ∈ Rd . Furthermore, denote the class of normalized order d
tensors with m dimensions in each mode and nonnegative elements by
m
X
Ad,m = {A ∈ Rm×···×m : Ai1 ,...,id ≥ 0,                                  Ai1 ,...,id = 1}.           (2)
i1 ,...,id =1

Definition 1 (Marginalizable Density Model Approximator). For Φ : Rd × [m]d → [0, 1] as above
and A ∈ Ad,m , the marginalizable density model approximator (MDMA) is
m
X                        d
Y
FA,Φ (x) = hA, Φ(x)i =                            Ai1 ,...,id         ϕij ,j (xj ).              (3)
i1 ,...,id =1                 j=1

If is clear that the conditions on Φ and A imply that FA,Φ ∈ Fd . As in the bivariate case, densities
or marginalization over xj are obtained by replacing each ϕi,j respectively by ϕ̇i,j or 1. As for

3```
```conditioning, considering any disjoint subsets R = {k1 , . . . , kr } and S = {j1 , . . . , js } of [d] such
that R ∩ S = ∅, we have
m
P                  Q               Q
Ai1 ,...,id        ϕik ,k (xk ) ϕ̇ij ,j (xj )
i1 ,...,id =1               k∈R                   j∈S
FA,Φ (xk1 , . . . , xkr |xj1 , . . . , xjs ) =                   m
P                       Q                     .
Ai1 ,...,id         ϕ̇ij ,j (xj )
i1 ,...,id =1                 j∈S

For a completely general tensor A with md parameters, the expression (3) is computationally
impractical, hence some structure must be imposed on A. For instance, one simple choice is
Pm      Qd
Ai1 ,...,id = ai1 δi1 ,...,id , which leads to FA,Φ (x) =  i=1 ai  j=1 ϕi,j (xj ), with ai ≥ 0 and
Pm
i=1 ai = 1. Instead of this top-down approach, A can be tractably constructed bottom-up, as we
explain next.
p−`+1
Assuming d = 2p for integer p, define ϕ` : Rd → Rm×2          for ` ∈ {1, . . . , p} recursively by

ϕi,j (xj ),                         `=1
ϕ`i,j (x) = P  m
`−1 `−1          `−1                                          (4)
     λi,k,j ϕk,2j−1 (x)ϕk,2j (x), ` = 2, . . . , p
k=1

Pmi ∈ [m],
for          j ∈ [2p−`+1 ], and where λ` is a non-negative m × m × 2p−`+1 tensor, normalized as
`
k=1 λi,k,j = 1. The joint CDF can then be written as
Xm
FAHT ,Φ (x) =     λpk ϕpk,1 (x)ϕpk,2 (x),                    (5)
k=1
Pm    p                                                     HT
with λp ∈ Rm+ satisfying  k=1 λk = 1. It is easy to verify that the underlying A         satisfies
HT
A ∈ Ad,m defined in (2). A graphical representation of this tensor is provided in Figure 10 in the
supplementary materials.
For example, for d = 4, we first combine (x1 , x2 ) and (x3 , x4 ) into
m
X                                               Xm
ϕ2i,1 (x) =   λ1i,k,1 ϕk,1 (x1 )ϕk,2 (x2 ) ,   ϕ2i,2 (x) =       λ1i,k,2 ϕk,3 (x3 )ϕk,4 (x4 ) ,
k=1                                                              k=1
and then merge them as
m
X
FAHT ,Φ (x) =              λ2k ϕ2k,1 (x)ϕ2k,2 (x),                                     (6)
k=1
Pm
from which we can read off that AHT
i1 ,i2 ,i3 ,i4 =               k=1   λ2k λ1k,i1 ,1 λ1k,i3 ,2 δi1 ,i2 δi3 ,i4 .
Note that the number of parameters required to represent AHT is only poly(m, d). The construction
is easily generalized to dimensions d not divisible by 2. Also, the number of ϕ factors combined at
each iteration in (4) (called the pool size), can be any positive integer. This construction is a variant
of the hierarchical Tucker decomposition of tensors , which has been used in the construction of
tensor networks for image classification in .
Given a set of training points {x}N
i=1 , we fit MDMA models by maximizing the log of the density with
respect to both the parameters in {ϕi,j }i∈[m],j∈[d] and the components of A. We present additional
details regarding the choice of architectures and initialization in Appendix E.

3.4   A non-marginalizable MDMA

We can construct a more expressive variant of MDMA at the price of losing the ability to marginalize
and condition on arbitrary subsets of variables. We find that the resulting model leads to state-of-
the-art performance on a density estimation benchmark. We define v = x + Tσ(x) where T is an
upper-triangular matrix with non-negative entries and 0 on the main diagonal. Note that ∂v ∂x = 1.
d
Q
Given some density f (v1 , . . . , vd ) =   ϕ̇j (vj ), we have
j=1
∂v
f (x1 , . . . , xd ) =      f (v1 , . . . , vd ) = f (v1 (x), . . . , vd (x)).                        (7)
∂x

4```
```We refer to this model nMDMA, since it no longer enables efficient marginalization and conditioning.

3.5   MDMA sampling

Given an MDMA as in (3), we can sample in the same manner as for autoregressive models: from
u1 , . . . , ud independent U (0, 1) variables, we obtain a sample from FA,Φ by computing
−1                  −1                         −1
x1 = FA,Φ (u1 ), x2 = FA,Φ    (u2 |x1 ) · · · xd = FA,Φ  (ud |x1 , . . . , xd−1 ),
where, unlike with autoregressive model, the order of taking the conditionals does not need to be
fixed. The main drawback of this method is that due to the sequential nature of the sampling the
computational cost is linear in d and cannot be parallelized.
However, the structure of FA,Φ can be leveraged to sample far more efficiently. Define by RA a
vector-valued categorical random variable taking values in [m] × · · · × [m], with distribution
P [RA = (i1 , . . . , id )] = Ai1 ,...,id .
The fact that A ∈ Ad,m with Ad,m from (2) ensure the validity of this definition. Consider a vector
(X̃1 , . . . , X̃d , RA ) where RA is distributed as above, and
h                                     i Y     d
P X̃1 ≤ x1 , . . . , X̃d ≤ xd |RA = r =              ϕri ,i (xi ),
i=1
for the collection {ϕi,j } of univariate CDFs. Denoting the distribution of this vector by F̃A,Φ ,
marginalizing over RA gives
X                d
Y
F̃A,Φ (x) =         P [RA = r]   ϕrj ,j (xj ) = FA,Φ (x).
r∈[m]d                 j=1

Instead of sampling directly from the distribution FA,Φ , we can thus sample from F̃A,Φ and discard
the sample of RA . To do this, we first sample from the categorical variable RA . Denoting this sample
by r, we can sample from X̃i by inverting the univariate CDF ϕri ,i . This can be parallelized over i.
The approach outlined above is impractical since RA can take md possible values, yet if A can be
expressed by an efficient tensor representation this exponential dependence can be avoided. Consider
the HT decomposition (5), which can be written as
p−`
p 2Y X
Y    m
HT
FAHT ,Φ (x) = A        , Φ(x) =                        λ`k`+1,                        Φ(x)k1,1 ,...,k1,d/2 ,   (8)
dj` /2e ,k`,j` ,j`
`=1 j` =1 k`,j` =1

that is a normalized sum of O(md) univariate CDFs.
Proposition 1. Sampling from (8) can be achieved in O(log d) time requiring the storage of only
O(d) integers.
Note that the time complexity of this sampling procedure depends only logarithmically on d. The
reason is that a simple hierarchical structure of Algorithm 1, where Mult denotes the multinomial
distribution.
Algorithm 1: Sampling from the HT MDMA
Result: xj for j = 1, . . . , d
kp,1 ∼ Mult(λp∗ );
for ` ← p − 1 to 1 by −1 do
k`,j ∼ Mult(λ`k`+1,dj/2e ,∗,j ),       j = 1, . . . , 2p−` ;
end
xj ∼ ϕk1,dj/2e ,j ,    j = 1, . . . , 2p ;

The logarithmic dependence is only in the sampling from the categorical variables, which is inexpen-
sive to begin with. We thus avoid the linear dependence of the time complexity on d that is common
in sampling from autoregressive models. Furthermore, the additional memory required for sampling
scales like log m (since storing the categorical samples requires representing integers up to size m),
and aside from this each sample requires evaluating a single product of univariate CDFs (which is
independent of m). In preliminary experiments, we have found that even for densities with d ≤ 10,
this sampling scheme is faster by 1.5 to 2 orders of magnitude than autoregressive sampling. The
relative speedup should only increase with d.

5```
```3.6        Universality of the MDMA

To model functions in F1 , we use Φl,r,σ , a class of constrained feedforward neural networks proposed
in  with l hidden layers, each with r neurons, and σ a nonaffine, increasing and continuously
differentiable elementwise activation function, defined as
Φl,r,σ = {ϕ : R → [0, 1], ϕ(x) = sigmoid ◦ Ll ◦ σ ◦ Ll−1 ◦ σ · · · ◦ σ ◦ L1 ◦ σ ◦ L0 (x)},
where Li : Rni → Rni+l is the affine map Li (x) = Wi x + bi for an ni+1 × ni weight matrix Wi
with nonnegative elements and an ni+1 × 1 bias vector bi , with nl+1 = n0 = 1 and ni = r for
i ∈ [l]. The constraints on the weights and the final sigmoid guarantee that Φl,r,σ ⊆ F1 , and for any
ϕ ∈ Φl,r,σ , the corresponding density ϕ̇(x) = ∂ϕ(x)/∂x can be obtained with the chain rule. The
universal approximation property of the class Φl,r,σ is expressed in the following proposition.
Proposition 2. ∪l,r Φl,r,σ is dense in F1 with respect to the uniform norm.
While the proof in the supplementary assumes that limx→−∞ σ(x) = 0 and limx→∞ σ(x) = 1,
it can be easily modified to cover other activations. For instance, in our experiments, we use
σ(x) = x + a tanh(x) following , and refer to the supplementary material for more details
regarding this case. In the multivariate case, consider the class of order d tensored-valued functions
with m dimensions per mode defined as
d
Φm,d,l,r,σ = {Φ : Rd × [m]d → [0, 1], Φ(x)i1 ,...,id =
Q
ϕij ,j (xj ), ϕi,j ∈ Φl,r,σ }.
j=1

Combining Φm,d,l,r,σ with the Ad,m , the normalized tensors introduced in Section 3.3, the class of
neural network-based MDMAs can then be expressed as
MDMAm,d,l,r,σ = {FA,Φ : Rd → [0, 1], FA,Φ (x) = hA, Φ(x)i, A ∈ Ad,m , Φ ∈ Φm,d,l,r,σ }.
Proposition 3. The set ∪m,l,r MDMAm,d,l,r,σ is dense in Fd with respect to the uniform norm.
The proof relies on the fact that setting m = 1 yields a class that is dense in the space of d-dimensional
CDFs with independent components. All proofs are provided in Appendix A.

4         Experiments
Additional experimental details for all experiments are provided in Appendix C.1

4.1        Toy density estimation

We start by considering 3D augmentations of three popular 2D toy probability distributions introduced
in : two spirals, a ring of 8 Gaussians and a checkerboard pattern. These distributions allow to
explore the ability of density models to capture challenging multimodalities and discontinuities [6, 9].
The results, presented in Figure 1, show that MDMA captures all marginal densities with high
accuracy, and samples from the learned model appear indistinguishable from the training data.

4.2        Mutual information estimation

Given a multivariate probability distribution over some variables X = (X1 . . . , Xd ), estimating the
mutual information                                                    
pX (x)
Z
I(Y ; Z) = dpX (x) log                     ,                         (9)
pY (y)pZ (z)
where Y, Z are random vectors defined by disjoint subsets of the Xi , requires evaluating pY , pZ
which are marginal densities of X. Typically, Y and Z must be fixed in advance, yet in some
cases it is beneficial to be able to flexibly compute mutual information between any two subsets of
variables. Estimating both I(Y, Z) and I(Y 0 , Z 0 ) may be highly inefficient, e.g. if Y and Y 0 are
highly overlapping subsets of X. Using MDMA however, we can fit a single model for the joint
distribution and easily estimate the mutual information between any subset of variables by simply
marginalizing over the remaining variables to obtain the required marginal densities. Thus a Monte
Carlo estimate of (9) can be obtained by evaluating the marginal densities at the points that make up
the training set. Figure 2 presents an example of this method, showing the accuracy of the estimates.
1
Code for reproducing all experiments is available at https://github.com/dargilboa/mdma.

6```
```Training data
Samples from model
x2

x3

x3

x3
x1

x1
x2
x2
x1                        x1
f (x1, x2)                f (x1, x3)             f (x1, x2|x3 = 0)             f (x1, x2|x3 = 0.5)

x2

x2
x2

x3

x1                        x1                               x1                          x1
f (x1, x2)                                               f (x1, x2)
Training data                                             Training data
x3
x3
x2

x2
x2
x1                x2                                                    x1

x1                                                           x1

Figure 1: Density estimation with closed-form marginals and conditionals. First Row: Panels 1,2:
Empirical histograms of training data. Panel 3: Samples from the training data. Panel 4: Samples
from the trained MDMA model. Second Row: Panels 1,2: The marginal density learned by MDMA
plotted on a grid. Panels 3,4: Conditional densities learned by MDMA plotted on a grid. Third
Row: Results on additional datasets: Panels 1,2: Training data and learned marginal density for a 3D
checkerboard dataset. Panels 3,4: Similarly for a 3D mixture of Gaussians.
I((X1, ..., Xk ); (Xk+1, ..., Xd))

Figure 2: Mutual information estimation be-
0.4                                                    tween subsets of a random vector. We fit-
ted a single MDMA model to samples from a
0.3                                                    zero-mean d = 16 Gaussian, with covariance
Σij = δij + (1 − δij )(i + j − 2)/(5d). Monte
0.2                                                    Carlo estimates of the mutual information (9)
between (X1 , . . . , Xk ) and (Xk+1 , . . . , Xd ) for
Ground Truth        any k = 1, . . . , d − 1 are easily obtained and
0.1                                                    match closely the exact values. For each k we
MDMA
average over 5 repetitions of drawing the dataset
1 2 3 4 5 6 7 8 9 10 11 12 13 14 15            and fitting.
k

4.3                                    Density estimation with missing values

Dealing with missing values in multivariate data is a classical challenge in statistics that has been
studied for decades . The standard solution is the application of a data imputation procedure
(i.e., “filling in the blanks”), which requires making structural assumptions. In some cases, this
is natural, as for the matrix completion problem under a low-rank assumption [40, 41], where the
imputed values are the main object of interest. But the artifacts introduced by data imputation  are
generally a price that one must unwillingly pay in order to perform statistical inference in models that

7```
```POWER dataset                                         GAS dataset
−0.2                                                            MDMA
MDMA

Test NLL
Test NLL
BNAF + MICE                                        BNAF + MICE
−5
−0.4

−0.6                                                      −10
0.1     0.2      0.3      0.4     0.5                    0.5       0.6       0.7      0.8
Missingness probability                             Missingness probability
Figure 3: Density estimation with missing data Test NLL on two density estimation benchmarks,
varying the proportion of entries in each data matrix that are designated missing and not used for
fitting. We compare MDMA which can fit marginal densities directly with BNAF which achieves
state-of-the-art results on the POWER dataset, after performing data imputation using MICE. As the
proportion of missing data increases, MDMA outperforms BNAF.

require fully-observed data points. Two popular, generic techniques for imputation are MICE 
and k-NN imputation . The former imputes missing values by iteratively regressing each missing
variable against the remaining variables, while the latter uses averages over the non-missing values at
k-nearest datapoints.
More formally, let X ∈ Rd be distributed according to some density p with parameters θ, let
X(0) , X(1) be the non-missing and missing entries of X respectively, and M ∈ {0, 1}d a vector
indicating the missing entries. In the missing-at-random setting (i.e. M is independent of X(1) ),
likelihood-based inference using the full likelihood
R         of the model is equivalent to inference using
the marginal likelihood  L(X(0) |θ) = p(X|θ)dX(1) . Standard neural network-based density
estimators must resort to data imputation because of the impossibility of computing this marginal
likelihood. MDMA however can directly maximize the marginal likelihood for any pattern of missing
data at the same (actually slightly cheaper) computational cost as maximizing the full likelihood,
without introducing any bias or variance due to imputation.
As a demonstration of this capability, we consider the UCI POWER and GAS datasets, following
the same pre-processing as . We construct a dataset with missing values by setting each entry in
the dataset to be missing independently with a fixed probability . We compare MDMA to BNAF ,
a neural density model which achieves state-of-the-art results on a number of density estimation
benchmarks including GAS. We train MDMA directly on the log marginal likelihood of the missing
data, and BNAF by first performing data imputation using MICE  and then training using the full
log likelihood with the imputed data. The validation loss is the log marginal likelihood for MDMA
and the log likelihood of the imputed validation set for BNAF. The test set is left unchanged for both
models and does not contain any missing values. We train BNAF using the settings specified in 
that led to the best performance (2 layers and 40d hidden units where d is the dimensionality of the
dataset). The results are shown in Figure 3. We find that, as the probability of missingness increases,
MDMA significantly outperforms BNAF on both datasets. Note that, while the proportion of missing
values might seem extreme, it is not uncommon in some applications (e.g., proteomics data). We
also trained BNAF using k-NN imputation , finding that performance was worse than MICE
imputation for all values of α. A comparison of the two methods is provided in Appendix B.

4.4          Conditional independence testing and causal discovery

Randomized control trials  remain the golden standard for causal discovery. Nonetheless,
experiments or interventions are seldom doable, e.g. due to financial or ethical considerations.
Alternatively, observational data can help uncovering causal relationships [47, 48]. In this context, a
class of popular methods targeted at recovering the full causal graph, like PC or FCI [47, 49], rely on
conditional independence (CI) tests. Letting X, Y and Z be random variables, the CI of X and Y
given Z, denoted X ⊥ ⊥ Y | Z, means that given Z, no information about X (or Y ) can be gained by
knowing the value of Y (or X). And testing H0 : X ⊥     ⊥ Y | Z against H1 : X 6⊥
⊥ Y | Z is a problem
tackled in econometrics [50, 51], statistics [52, 53], and machine learning [54, 55].
Following , denote U1 = F (X | Z) and U2 = F (Y | Z). It is clear that H0 implies U1 ⊥       ⊥ U2 ,
although the converse does not hold [see e.g., 56]. Nonetheless, U1 6⊥
⊥ U2 implies H1 , so a test based

8```
```Table 1: MDMA for causal discovery. Conditional densities from a trained MDMA model can be
used for causal discovery by allowing to test for conditional independence between variables. Both
on synthetic DAG data and real data from a protein signaling network, MDMA infers the graph
structure more accurately than a competing method based on quantile regression . The metrics
are the structural Hamming distance for the directed (SHD(D)) and undirected (SHD) graph.
Model         Sigmoidal DAG, d=10        Polynomial DAG, d=10            Sachs , d=11
SHD(D)         SHD          SHD(D)          SHD         SHD(D)          SHD
Gaussian 18.6 ± 3.0 15.6 ± 2.7 19.8 ± 4.1 18.9 ± 4.2                     32            27
MDMA         15.6 ± 6.1 12.8 ± 5.2 17.9 ± 5.3 15.0 ± 4.5 30.3 ± 1.8 25.8 ± 0.7

on the independence between U1 and U2 can still have power. While the test from  is based on
estimating the conditional CDFs through quantile regression, we proceed similarly, albeit using the
MDMA as a plugin for the conditional distributions. Our approach is especially appealing in the
context of causal discovery, where algorithms require computing many CI tests to create the graph’s
skeleton. Instead of having to regress for every test, MDMA estimates the full joint distribution, and
its lower dimensional conditionals are then used for the CI tests.
In Table 1, we present results on inferring the structure of causal graphs using the PC algorithm
[47, 57–60]. As a benchmark, we use the vanilla (i.e., Gaussian) CI test, and compare it to the PC
algorithm obtained with the CI test from , albeit using the MDMA for the conditional distributions.
Synthetic random directed acyclic graphs (DAGs) along with sigmoidal or polynomial mechanisms
linking parents to children are sampled using . Each dataset is d = 10 dimensional and contains
20,000 observations. We also compare the two algorithms on data from a protein signaling network
with d = 11  for which the ground truth causality graph is known. Performance is assessed based
on the structural Hamming distance (SHD) , that is the L1 norm of the difference between learned
adjacency matrices and the truth, as well as a variant of this metric for directed graphs SHD(D) which
also accounts for the direction of the edges. Table 1 shows averages over 8 runs for each setting. In
all cases, MDMA outperforms the vanilla PC in terms of both metrics. For the synthetic data, we
note the large standard deviations, due in part to the fact that we sample randomly from the space of
DAG structures, which has cardinality super-exponential in d. An example of the inferred graphs is
presented in Appendix B.

4.5        Density estimation on real data

We trained MDMA/nMDMA and the non-marginalizable variant described in Section 3.4 on a number
of standard density estimation benchmarks from the UCI repository,2 following the pre-processing
described in . Table 2 compares test log likelihoods of MDMA/nMDMA with several other neural
density models. We find the performance of MDMA on the lower-dimensional datasets comparable to
state-of-the-art models, while for higher-dimensional datasets it appears to overfit. nMDMA achieves
state-of-the-art performance on the POWER (d = 6) dataset, but at the cost of losing the ability to
marginalize or condition over subsets of the variables. The width of MDMA was chosen based on a
grid search over {500, 1000, 2000, 3000, 4000} for each dataset, and the marginal CDF parameters by
a search over {(l = 2, w = 3) , (l = 4, w = 5)}. All models were trained using ADAM with learning
rate 0.01, and results for MDMA and nMDMA are averaged over 3 runs. Additional experimental
details are provided in Appendix C.

5         Discussion
MDMAs offer the ability to obtain, from a single model, closed form probabilities, marginals and
conditionals for any subset of the variables. These properties enable one to straightforwardly use
the model to solve a diverse array of problems, of which we have demonstrated only a few: mutual
information estimation between arbitrary subsets of variables, inference with missing values, and
conditional independence testing targeted at multivariate causal discovery. In addition to these,
MDMA’s marginalization property can be used for anomaly detection with missing values . We
have shown that MDMA can fit data with missing values without requiring imputation, yet if one is

2
http://archive.ics.uci.edu/ml/datasets.php

9```
```Table 2: General density estimation. Test log likelihood for density estimation on UCI datasets.
The comparison results are reproduced from .
Model                       POWER [d=6]     GAS [d=11]    HEPMASS [d=21] MINIBOONE [d=43]
Kingma et al. 2018         0.17 ± .01      8.15 ± .4     −18.92 ± .08         −11.35 ± .07
Grathwohl et al. 2019      0.46 ± .01     8.59 ± .12     −14.92 ± .08         −10.43 ± .04
Huang et al. 2018          0.62 ± .01    11.96 ± .33      −15.08 ± .4          −8.86 ± .15
Oliva et al. 2018          0.60 ± .01    12.06 ± .02     −13.78 ± .02         −11.01 ± .48
De Cao et al. 2019         0.61 ± .01    12.06 ± .09     −14.71 ± .38          −8.95 ± .07
Bigdeli et al. 2020       0.97 ± .01    9.73 ± 1.14      −11.3 ± .16         −6.94 ± 1.81
MDMA                          0.57 ± .01    8.92 ± 0.11      −20.8 ± .06          −29.0 ± .06
nMDMA                        1.78 ± .12      8.43 ± .04     −18.0 ± 0.91          −18.6 ± .47

interested in data imputation for downstream tasks, the ability to sample from arbitrary conditional
distributions means that MDMA can be used for imputation as well.
Additionally, in some application areas (e.g., financial risk management), powerful models exist for
the univariate distributions, and marginal distributions are then glued together using copulas .
However, popular copula estimators suffer from the same drawbacks as modern neural network density
estimators with regard to marginalization and conditioning. Using MDMA for copula estimation (say
by replacing the kernel density estimator by MDMA in the formulation of ), one can then obtain
copula estimators that do not suffer from these deficiencies.
The main shortcoming of MDMA is the linearity in the combination of the products of univariate
CDFs which appears to limit the expressivity of the model. The study of tensor decompositions is an
to improvements in this regard despite the linear structure.

10```
```Acknowledgements
The work of DG is supported by a Swartz fellowship. The work of AP is supported by the Simons
Foundation, the DARPA NESD program, NSF NeuroNex Award DBI1707398 and The Gatsby
Charitable Foundation.

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14```
```Supplementary Material
A     Proofs
A.1     Proof of Proposition 1

The proof follows directly from Algorithm 1. The distribution FAHT ,Φ ) = AHT , Φ is a mixture
model, and thus in order to sample from it we can first draw a single mixture component (which is a
product of univariate CDFs) and then sample from this single component. The mixture weights are
the elements of the tensor AHT given by the diagonal HT decomposition (8). In the next section, we
add details on the sampling process for the sake of clarity.

A.1.1    Details on the sampling for the HT model
`
Define a collection of independent categorical variables R = {Ri,j    } taking values in [m], where
p−`
` ∈ [p], i ∈ [m] and for any `, j ∈ [2 ]. These variables are distributed according to
 `
= k = λ`i,k,j ,

∀`, i, j : P Ri,j

where {λ` }p`=1 are the parameters of the HT decomposition. The fact that the parameters are
Pm
nonnegative and k=1 λ`k,i,j = 1 ensures the validity of this distribution.
With the convention Rkpp+1,1 ,1 = R1,1
p
, define the event
p            n                                   o
\         p−`                                            p
∩2j` =1 Rk` `+1,                 = k`,j`        = R1,1 = kp,1
dj` /2e ,j`
`=1
\           n                    o
∩2j=1 Rkp−1
p,1 ,j
= k p−1,j

..
.
\       d/2
n                    o
∩j=1 Rk12,dj/2e ,j = k1,j    .

Let (X̃1 , . . . , X̃d , R) be a random vector such that
p 
"                                                                                            #       d
\       p−`
n                                          o           Y
P X̃1 ≤ x1 , . . . , X̃d ≤ xd           ∩2j` =1 Rk` `+1,                        = k`,j`            =         ϕk1,di/2e ,i (xi ), (10)
dj` /2e ,j`
`=1                                                                   i=1

which implies that the distribution of (X̃1 , . . . , X̃d ) obtained after conditioning on a subset of the
`
{Ri,j } in this way is equal to a single mixture component in FHT = hA, Φi. Thus, based on
a sample of R, one can sample X̃i by inverting the univariate CDFs ϕk1,di/2e ,i numerically and
parallelizing over i. Numerical inversion is trivial since the functions are increasing and continuously
differentiable, and this can be done for instance using the bisection method. It remains to sample a
mixture component.
`
Assume that a sample {Ri,j } for a sequence of variables as in (10) is obtained e.g. from Algorithm 1.
With the convention λkp+1,1 ,k,1 = λpk , since
p

p−`
p                                                                p 2Y
"                                                         #
\     p−`
n                                          o           Y
P    ∩2j` =1 Rk` `+1,                        = k`,j`            =               λ`k`+1,                        ,
dj` /2e ,j`                                                dj` /2e ,k`,j` ,j`
`=1                                                           `=1 j` =1

sampling from the categorical variables in this fashion is equivalent to sampling a mixture component.
It follows that by first sampling a single mixture component and then sampling from this component,
one obtains a sample from FHT .
The main loop in Algorithm 1 samples such a mixture component, and there are p = log2 d layers in
the decomposition, so the time complexity of the main loop is O(log d), and aside from storing the
decomposition itself this sampling procedure requires storing only O(d) integers. This logarithmic

15```
```dependence is only in sampling from the categorical variables which is computationally cheap. This
not only avoids the linear time complexity common in sampling from autoregressive models (without
using distillation), but the space complexity is also essentially independent of m since only a single
mixture component is evaluated per sample.

A.2     Proof of Proposition 2

Assume that the activation function σ is increasing, continuously differentiable, and such that
limx→−∞ σ(x) = 0 and limx→∞ σ(x) = 1. Proposition 2 then follows immediately from Proposi-
tion 4 and the fact that ∪r Φ1,r,σ ⊆ ∪l,r Φl,r,σ .
Remark 1. In practice, we use the activation σ(x) = x + a tanh(x) for some a > −1. While it
does not satisfy the assumptions, the arguments in the proof of Proposition 5 can be modified in a
straightforward manner to cover this activation (see Remark 2).
Proposition 4. ∪r Φ1,r,σ is dense in F1 with respect to the uniform norm.

Letting
Fe1 = {Fe : R → R, Fe(x) = log F (x)/(1 − F (x)), F ∈ F1 },
Φ
e l,r,σ = {ϕ
e : R → R, ϕ(x)
e    = log ϕ(x)/(1 − ϕ(x)), ϕ ∈ Φl,r,σ },
the proof of Proposition 4 relies on the following proposition.
Proposition 5. ∪r Φe 1,r,σ is dense in Fe1 with respect to the uniform norm.

A.2.1     Proof of Proposition 4

This proof is similar to that of [67, Theorem 2], which deals with functions with positive outputs. We
want to show that, for any F ∈ F1 , compact K ⊂ R, and  > 0, there exists ϕ ∈ ∪r Φ1,r,σ such that
kϕ − F k∞,K = sup |ϕ(x) − F (x)| ≤ .
x∈K

Denote the sigmoid function by ρ(x) = 1/(1 + e−x ) and define the function Fe : R → R by
Fe(x) = log F (x)/(1 − F (x)), so that F = ρ ◦ Fe. By Proposition 5, there exists ϕ
e ∈ ∪r Φ
e 1,r,σ such
that
sup |ϕ(x)
e    − Fe(x)| ≤ 4.
x∈K

Thus, letting ϕ = ρ ◦ ϕ,
e we have
|ϕ(x) − F (x)| = |ρ ◦ ϕ(x)
e    − ρ ◦ Fe(x)| ≤ sup ρ(x)0 |ϕ(x)
e    − Fe(x)| ≤ .
x∈K

Since Φ1,r,σ = {ρ ◦ ϕ
e: ϕ
e∈Φ
e 1,r,σ }, the result follows.

A.2.2     Proof of Proposition 5

This proof is similar to that of [68, Theorem 3.1], which is incomplete and only deals with the sigmoid
activation. First, note that Fe1 is the space of strictly increasing and continuously differentiable
functions. Therefore, for any Fe ∈ Fe1 and interval K = [K1 , K2 ], we can write, for any x ∈ K,
Z Fe(K2 )                         Z Fe(K2 )
F (x) = F (K1 ) +
e         e                    1Fe(x)≥u du = F (K1 ) +
e                   1x≥Fe−1 (u) du,
e(K1 )
F                                e(K1 )
F

where the existence of the inverse Fe−1 is guaranteed by the fact that Fe is strictly increasing and
continuous. Thus, for Fe(K1 ) = u0 < u1 < · · · < uk = Fe(K2 ) a partition of [Fe(K1 ), Fe(K2 )] with
uj+1 − uj ≤ /2(Fe(K2 ) − Fe(K1 )), xj = Fe−1 (uj ) and
k
X
G(x) = Fe(K1 ) +          1x≥xj (uj − uj−1 ),
j=1

16```
```we have |Fe(x) − G(x)| ≤ /2, namely the approximation error of the Riemann sum for increasing
functions. Let a > 0 and ϕ  e ∈ Φ
e 1,k,σ obtained by setting b1 = Fe(K1 ), as well as (W1 )1,j =
(uj − uj−1 )/a > 0, (b0 )j = −axj and (W0 )j,1 = a for 1 ≤ j ≤ k, then
k
X
|G(x) − ϕ(x)|
e     ≤               (uj+1 − uj ) 1x≥xj − σ(a(x − xj )) .
j=1
By the assumptions on σ, it is clear that |1x≥0 − σ(ax)| can be made arbitrarily small. Thus, taking
a large enough so that |G(x) − ϕ(x)|
e      ≤ /2, we have
|Fe(x) − ϕ(x)|
e      ≤ |Fe(x) − G(x)| + |G(x) − ϕ(x)|
e     ≤ .
Remark 2. Let σ(x) = x + a tanh(x) for some a > −1 and ϕ        e ∈ Φ e 1,k,σ obtained by setting
b1 = F (K1 ) + 1/2, as well as (W1 )1,j = (uj − uj−1 )/2a > 0, (b0 )j = −axj and (W0 )j,1 = |a|
e
for 1 ≤ j ≤ k, then
X k
|G(x) − ϕ(x)|
e     ≤      (uj+1 − uj ) 1x≥xj − tanh(|a|(x − xj ))/2 − 1/2
j=1
k
X
+         (uj+1 − uj )|x − xj |/2|a|.
j=1
Because a is arbitrary, one can take it large enough so that |G(x) − ϕ(x)|
e     ≤ /2 as above.

A.3     Proof of Proposition 3

Consider the classes of order d tensored-valued functions with m dimensions per mode defined as
d
Φm,d,l,r,σ = {Φ : Rd × [m]d → [0, 1], Φ(x)i1 ,...,id =
Q
ϕij ,j (xj ), ϕi,j ∈ Φl,r,σ },
j=1
d
Fm,d = {Φ : Rd × [m]d → [0, 1], Φ(x)i1 ,...,id =
Q
Fij ,j (xj ), Fi,j ∈ F1 }
j=1
as well as the class of neural network-based and F1 -based MDMAs, that is
MDMAm,d,l,r,σ = {FA,Φ : Rd → [0, 1], FA,Φ (x) = hA, Φ(x)i, A ∈ Ad,m , Φ ∈ Φm,d,l,r,σ },
MDMAm,d,F1 = {FA,Φ : Rd → [0, 1], FA,Φ (x) = hA, Φ(x)i, A ∈ Ad,m , Φ ∈ Fm,d }.
We can now state the following proposition.
Proposition 6. ∪l,r MDMAm,d,l,r,σ is dense in MDMAm,d,F1 with respect to the uniform norm
Proposition 3 then follows immediately from the fact that ∪m MDMAm,d,F1 is the space of multivari-
ate mixture distributions admitting a density, which is dense in Fd with respect to the uniform norm
(see e.g., [69, Theorem 33.2], [70, Theorem 5], or [71, Corollary 11]).

A.3.1    Proof of Proposition 6

With A ∈ Ad,m , Φ1 ∈ Φm,d,l,r,σ , and  > 0 and a compact K = K1 × · · · × Kd ⊂ Rd , we want to
prove that there exists A2 ∈ Ad,m and Φ2 ∈ Fm,d , such that supx∈K |FA,Φ1 (x) − FA2 ,Φ2 (x)| ≤ .
Assuming that we can show supx∈K |Φ1 (x)i1 ,...,id − Φ2 (x)i1 ,...,id | ≤ , the result would then follow
from setting A2 = A and the fact that FA,Φ1 (x) − FA,Φ2 (x) = hA, Φ1 (x) − Φ2 (x)i implies
X
sup |FA,Φ1 (x) − FA,Φ2 (x)| = ≤          Ai1 ,...,id sup |Φ1 (x)i1 ,...,id − Φ2 (x)i1 ,...,id | = .
x∈K                                     i1 ,...,id             x∈K

With δ = 1/d , by Proposition 2, there exists l, w, and {ϕ}i∈[m],j∈[d] with ϕi,j ∈ Φl,r,σ , such that
max         sup |ϕi,j (xj ) − Fi,j (xj )| ≤ δ.
i∈[m],j∈[d] xj ∈Kj

Thus, we have that
d
Y                      d
Y
|Φ1 (x)i1 ,...,id − Φ2 (x)i1 ,...,id | = |          ϕij ,j (xj ) −         Fi,j (xj )| ≤ δ d = .
j=1                    j=1

17```
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