The discretization trick

I explain the discretization trick that I mentioned in my previous post (Posterior consistency under possible misspecification). I think it was introduced by Walker (New approaches to Bayesian consistency (2004)).

Let \mathbb{F} be a set of densities and let \Pi be a prior on \mathbb{F}. If x_1, x_2, x_3, \dots follows some distribution P_0 having density f_0, then the posterior distribution of \Pi can be written as

\Pi(A | x_1, \dots, x_n) \propto \int_A \prod_{i=1}^n f(x_i) \Pi(df).

The discretization trick is to find densities f_{1}, f_2, f_3, \dots in the convex hull of A (taken in the space of all densities) such that

\int_A \prod_{i=1}^n f(x_i) = \prod_{i=1}^n f_i(x_i) \Pi(A).

For example, suppose \varepsilon > 2\delta > 0, A = A_\varepsilon = \{f \,|\, D_{\frac{1}{2}}(f_0, f) > \varepsilon\} and that A_i is a partition of A of diameter at most \delta. If there exists 1 > \alpha > 0 such that

\sum_i \Pi(A_i)^\alpha < \infty,

then for some \beta > 0 we have that

e^{n \beta} \left(\int_{A_{\varepsilon}} \prod_{i=1}^n \frac{f(x_i)}{f_0(x_i)} \Pi (df) \right)^\alpha \le e^{n \beta} \sum_i \prod_{j=1}^n \left(\frac{f_{i,j}(x_j)}{f_0(x_j)}\right)^\alpha \Pi(A_i)^\alpha  \rightarrow 0

almost surely. This is because, with A_{\alpha} the \alpha-affinity defined here, we have that


goes exponentially fast towards 0 when \beta is sufficiently small.  Hence the Borel-Cantelli lemma applies.


The f_{i,j}‘s are defined as the posterior mean predictive density, when the posterior is conditioned on A_i. That is,

f_{i,j} : x \mapsto \frac{\int_{A_i} f(x) \prod_{k=1}^{j-1}f(x_k) \Pi(df)}{\int_{A_i} \prod_{k=1}^{j-1}f(x_k) \Pi(df)}


f_{i, 1} : x \mapsto \frac{\int_{A_i} f(x) \Pi(df)}{\Pi(A_i)}.


\int_{A_i} \prod_{i=1}^n f(x_i) \Pi(df) = \prod_{j=1}^n f_{i,j}(x_j) \Pi(A_i).

Furthermore, if A_i is contained in a Hellinger ball of center g_i and of radius \delta, then also

H(f_{i,j}, g_i) < \delta.

This follows form the convexity of the Hellinger balls (an important remark for the generalization of this trick).

Posterior consistency under (possible) misspecification

We assume, without too much loss of generality, that our priors are discrete. When dealing with Hellinger separable density spaces, it is possible to discretize posterior distributions to study consistency (see this post about it).

Let \Pi be a prior on a countable space \mathcal{N} = \{f_1, f_2, f_3, \dots\} of probability density functions, with \Pi(f) > 0 for all f \in \mathcal{N}. Data X_1, X_2, X_3, \dots follows (independently) some unknown distribution P_0 with density f_0.

We denote by D_{KL}(f_0, f) = \int f_0 \log\frac{f_0}{f} the Kullback-Leibler divergence and we let D_{\frac{1}{2}}(f_0, f) = 1 - \int \sqrt{f_0 f} be half of the squared Hellinger distance.

The following theorem states that the posterior distribution of \Pi accumulates in Hellinger neighborhoods of f_0, assuming the prior is root-summable (i.e. \sum_{f \in \mathcal{N}} \Pi(f)^\alpha < \infty for some \alpha > 0) . In the well-specified case (i.e. \inf_{f \in \mathcal{N}} D_{KL}(f_0, f) = 0), the posterior accumulates in any neighborhood of f_0. In the misspecified case, small neighborhoods of f_0 could be empty, but the posterior distribution still accumulates in sufficiently large neighborhoods (how large exactly is a function of \alpha and \inf_{f \in \mathcal{N}} D_{KL}(f_0, f)).

The result was essentially stated by Barron (Discussion: On the Consistency of Bayes Estimates, 1986). In the case where \Pi is not necessarily discrete, a similar result was obtained, through a discretization argument, by Walker (Bayesian asymptotics with misspecified models, 2013). See also Xing (Sufficient conditions for Bayesian consistency, 2009) for a thorough treatment of Bayesian consistency using the same method of proof.

Theorem (Barron).
Suppose \beta_0 :=\inf_{f \in \mathcal{N}} D_{KL}(f_0, f) < \infty and that

\alpha := \inf \left\{ p \in [\tfrac{1}{2},1] \,|\, \sum_{f \in \mathcal{N}} \Pi(f)^p < \infty \right\} < 1.

If \varepsilon > 0 is such that

\varepsilon > 1- \exp\left( \frac{-\beta_0 \alpha}{2(1-\alpha)} \right)

and if A_\varepsilon := \{f \in \mathcal{N} \,|\, D_{\frac{1}{2}} (f_0, f) < \varepsilon\} \not = \emptyset, then

\Pi\left(A_\varepsilon \,|\, \{x_i\}_{i=1}^n\right) \rightarrow 1

almost surely as n \rightarrow \infty.


1 – If \inf_{f \in \mathcal{N}} D_{KL}(f_0, f) = 0, then any \varepsilon > 0 can be used.

2 – \alpha is related to the rate of convergence of \rho(n) := \Pi(f_n) towards 0. The quantity H_\alpha (\Pi) = \log \sum_{f \in \mathcal{N}} \Pi(f)^\alpha can be thought as measure of entropy.

3 – Walker (2013) considered the case \sum_{f \in \mathcal{N}} \Pi(f)^\alpha < \infty for some \alpha < \frac{1}{2}. This implies that \sum_{f \in \mathcal{N}} \sqrt{\Pi(f)} < \infty and the above theorem can also be applied in this case.


The proof is brief. I do not dwell on explanations.

First let me recall some concepts. The \alpha-affinity between two densities f and g is defined as

A_\alpha(f, g) = \int g^\alpha f^{1-\alpha}. \qquad (1)

Note that 0 \le A_{\alpha}(f, g) \le 1 and that A_\alpha(f, g) = 1 if and only if f = g. Furthermore, when \alpha \geq \frac{1}{2}, Holder’s inequality and Jensen’s inequality yield

A_{\frac{1}{2}}(f, g) \le A_{\alpha}(f, g) \le \left(A_{\frac{1}{2}}(f,g)\right)^{2(1-\alpha)}. \qquad (2)

We can now begin the proof. Let \tilde{\alpha} be such that 1 > \tilde{\alpha} > \alpha. Then, we have


If \beta > 0 and g \in \mathcal{N} is such that D_{KL}(f_0, g) < \beta_0 + \beta, then


almost surely. Furthermore, using (2), we find


Here \text{cst.} is a constant. Since \beta > 0 is arbitrary and since \tilde \alpha can be taken so that 2(1-\tilde \alpha) \log (1-\varepsilon) + \tilde \alpha \beta_0 < 0, we obtain that (**) converges exponentially fast towards 0. Hence, by the Borel-Cantelli lemma, we have


almost surely. This, together with (3), implies that (*) \rightarrow 0 almost surely. \Box


The choice of prior in bayesian nonparametrics – part 2

See part 1. Most proofs are omitted; I’ll post them with the complete pdf later this week.

The structure of \mathcal{M}

Recall that \mathbb{M} is is a Polish space (ie. a complete and separable metric space). It is endowed with its borel \sigma-algebra \mathfrak{B} which is the smallest family of subsets of \mathbb{M} that contains its topology and that is closed under countable unions and intersections. All subsets of \mathbb{M} we consider in the following are supposed to be part of \mathfrak{B}. A probability measure on \mathbb{M} is a function \mu : \mathfrak{B} \rightarrow [0,1] such that for any countable partition A_1, A_2, A_3, \dots of \mathbb{M} we have that \sum_{i=1}^\infty \mu(A_i) = 1. The set \mathcal{M} consists of all such probability measures.

Note that since \mathbb{M} is complete and separable, every probability measure \mu \in \mathcal{M} is regular (and tight). It means that the measure of any A\subset \mathbb{M} can be well approximated from the measure of compact subsets of A as well as from the measure of open super-sets of A:

\mu(A) = \sup \left\{\mu(K) \,|\, K \subset A \text{ is compact}\right\}\\ = \inf \left\{\mu(U) \,|\, U \supset A \text{ is open}\right\}.

Metrics on \mathcal{M}

Let me review some facts. A natural metric used to compare the mass allocation of two measures \mu, \nu \in \mathbb{M} is the total variation distance defined by

\|\mu - \nu\|_{TV} = \sup_{A \subset \mathbb{M}}|\mu(A) - \nu(A)|.

It is relatively straightforward to verify that \mathcal{M} is complete under this distance, but it is not in general separable. To see this, suppose that \mathbb{M} = [0,1]. If a ball centered at \mu contains a dirac measure \delta_x, x \in [0,1], then \mu must have a point mass at x. Yet any measure contains at most a countable number of point masses, and there is an uncountable number of dirac measures on [0,1]. Thus no countable subset of \mathcal{M} can cover \mathcal{M} up to an \varepsilon of error.

This distance can be relaxed to the Prokhorov metric, comparing mass allocation up to \varepsilon-neighborhoods. It is defined as

d_P(\mu, \nu) = \inf \left\{ \varepsilon > 0 \,|\, \mu(A) \le \nu(A^{\varepsilon}) + \varepsilon \text{ and } \nu(A) \le \mu(A^{\varepsilon}) + \varepsilon,\; \forall A \subset \mathbb{M} \right\},

where A^{\varepsilon} = \{x \in \mathbb{M} \,|\, d(x, A) < \varepsilon\} is the \varepsilon-neighborhood of A. It is a metrization of the topology of weak convergence of probability measures, and \mathcal{M} is separable under this distance.

The compact sets of \mathcal{M} under the Prokhorov metric admit a simple characterization given by the Prokhorov theorem: P \subset \mathcal{M} is precompact if and only if P is uniformly tight (for each \varepsilon > 0, there exists a compact K \subset X such that \sup_{\mu \in P} \mu(K) \geq 1-\varepsilon). This means that a sequence \{\mu_n\} \subset \mathcal{M} admits a weakly convergent subsequence if and only if \{\mu_n\} is uniformly tight.

Characterizations of weak convergence are given by the Portemanteau theorem, which says in particular that \mu_n converges weakly to \mu if and only if

\int f d\mu_n \rightarrow \int f d\mu

for all continuous and bounded and continuous f. It is also equivalent to

\mu_n(A) \rightarrow \mu(A)

for all sets A such that \mu(\partial A) = 0.

Measures of divergence

In addition to metrics, that admit a geometric interpretation through the triangle inequality, statistical measures of divergence can also be considered. Here, we consider functions D : \mathcal{M}\times \mathcal{M} \rightarrow [0, \infty] that can be used to determine the rate of convergence of the likelihood ratio

\prod_{i=1}^n \frac{d\mu}{d\nu}(x_i) \rightarrow 0,

where x_i \sim \nu and \mu, \nu \in \mathcal{M}.

Kullback-Leibler divergence

The weight of evidence in favor of the hypothesis “\lambda = \mu” versus “\lambda = \nu” given a sample x is defined as

W(x) = \log\frac{d\mu}{d\nu}.

It measures how much information about the hypotheses is brought by the observation of x. (For a justification of this interpretation, see Good (Weight of evidence: a brief survey, 1985).) The Kullback-Leibler divergence D_{KL} between \mu and \nu is defined as the expected weight of evidence given that x \sim \mu:

D_{KL}(\mu, \nu) =\mathbb{E}_{x \sim \mu} W(x) = \int \log \frac{d\mu}{d\nu} d\mu.

The following properties of the Kullback-Leibler divergence support its interpretation as an expected weight of evidence.

Theorem 1 (Kullback and Leibler, 1951).
We have

  1. D_{KL}(\mu, \nu) \geq 0 with equality if and only if \mu = \nu;
  2. D_{KL}(\mu T^{-1}, \nu T^{-1}) \geq D_{KL}(\mu, \nu) with equality if and only if T: \mathbb{M} \rightarrow \mathbb{M}' is a sufficient statistic for \{\mu, \nu\}.

Furthermore, the KL divergence can be used to precisely identify exponential rates of convergence of the likelihood ratio. The first part of the next proposition says that D_{KL}(\lambda, \nu) is finite if and only if the likelihood ratio \prod_{i} \frac{d\nu}{d\lambda}(x_i), x_i \sim \lambda cannot convergence super-exponentially fast towards 0. The second part identifies the rate of convergence then the KL divergence is finite.

Proposition 2.
Let x_1, x_2, x_3, \dots \sim \lambda (independently). The KL divergence D_{KL}(\lambda, \nu) is finite if and only if there exists an \alpha > 0 such that

e^{n\alpha} \prod_{i=1}^n \frac{d\nu}{d\lambda}(x_i) \rightarrow \infty

with positive probability.

Finally, suppose we are dealing with a submodel \mathcal{F} \subset \mathcal{M} such that the rates of convergences of the likelihood ratios in \mathcal{F} are of an exponential order. By the previous proposition, this is equivalent to the fact that \forall \mu, \nu \in \mathcal{F}, D_{KL}(\mu, \nu) < \infty. We can show that the KL divergence is, up to topological equivalence, the best measure of divergence that determines the convergence of the likelihood ratio. That is, suppose D: \mathcal{F} \times \mathcal{F}\rightarrow [0, \infty] is such that

D(\lambda, \mu) < D(\lambda, \nu) \Longrightarrow \prod_{i=1}^n \frac{d\nu}{d\mu}(x_i) \rightarrow 0

at an exponential rate, almost surely when x_i \sim \lambda, and that D(\lambda, \mu) = 0 if and only if \lambda = \mu. Then, the topology induced by D_{KL} is coarser than the topology induced by D.

Proposition 3.
Let D be as above and let \mathcal{F} \subset \mathcal{M} be such that \forall \mu, \nu \in \mathcal{F}, D_{KL}(\mu, \nu) < \infty. Then, the topology on \mathcal{F} induced by D_{KL} is weaker than the topology induced by D. More precisely, we have that

D(\lambda, \mu) < D(\lambda, \nu) \Rightarrow D_{KL}(\lambda, \mu) < D_{KL}(\lambda, \nu).


alpha-affinity and alpha-divergence

We define the \alpha-affinity between two probability measures as the expectancy of another transform of the likelihood ratio. Let \mu, \nu be two probability measures dominated by \lambda, with d\mu = f d\lambda and d\nu = g d\lambda. Given 0 < \alpha < 1, the \alpha-affinity between \mu and \nu is

A_\alpha(\mu, \nu) = \int \left(\frac{g}{f}\right)^\alpha d\mu = \int g^\alpha f^{1-\alpha} d\lambda.

Proposition 4.
For all 0 < \alpha < 1, we have that

1. A_\alpha(\mu, \nu) \le 1 with equality if and only if \mu = \nu;

2. A_\alpha is monotonous in \alpha and jointly concave in its arguments;

3. A_\alpha is jointly multiplicative under products:

A_\alpha (\mu^{(n)}, \nu^{(n)}) = \left(A_{\alpha}(\mu, \nu)\right)^n.

4. if \frac{1}{2} \leq \alpha, then

A_{\frac{1}{2}} \le A_\alpha \le \left(A_{\frac{1}{2}}\right)^{2(1-\alpha)};

1-2 follow from Jensen’s inequality and the joint concavity of (x,y) \mapsto x^\alpha y^{1-\alpha}. 3 follows from Fubini’s theorem. For
(iv), the first inequality is a particular case of 2 and Hölder’s inequality finally yields

A_{\alpha}(\mu, \nu) = \int (fg)^{1-\alpha} g^{2\alpha - 1} d\lambda \le \left( \int \sqrt{fg} \,d\lambda \right)^{2-2\alpha} = A_{\frac{1}{2}}(\mu, \nu).


The \alpha-divergence D_\alpha is obtained as

D_\alpha = 1 - A_\alpha.

Other similar divergences considered in the litterature are

\frac{1-A_\alpha}{\alpha(1-\alpha)}\; \text{ and }\; \frac{\log A_\alpha}{\alpha(1-\alpha)},

but we prefer D_\alpha for its simplicity. When \alpha = \frac{1}{2}, it is closely related to the hellinger distance

H(\mu, \nu) = \left(\int \left(\sqrt{f} - \sqrt{g}\right)^2d\lambda\right)^{\frac{1}{2}}


D_{\frac{1}{2}}(\mu, \nu) = \frac{H(\mu, \nu)^2}{2}.

Other important and well-known inequalities are given below.

Proposition 5.
We have

D_{\frac{1}{2}}(\mu, \nu) \le \|\mu-\nu\|_{TV} \le \sqrt{2 D_{\frac{1}{2}}(\mu, \nu)}


2D_{\frac{1}{2}}(\mu, \nu) \le D_{KL}(\mu, \nu) \le 2\left\|\frac{f}{g}\right\|_\infty \|\mu-\nu\|_{TV}.

This, together with proposition 4 (4) , yields similar bounds for the other divergences.

Finite models

Let \Pi be a prior on \mathcal{M} that is finitely supported. That is, \Pi = \sum_{i=1}^n p_i \delta_{\mu_i} for some \mu_i \in \mathcal{M} and p_i > 0 with \sum_i p_i = 1. Suppose that x_1, x_2, x_3, \dots independently follow some \mu_* \in \mathcal{M}.

The following proposition ensures that as data is gathered, the posterior distribution of \Pi concentrates on the measures \mu_i that are closest to \mu_*.

Proposition 6.
Let A_{\varepsilon} = \{\mu_i \,|\, D_{KL}(\mu_*, \mu_i) < \varepsilon \}. If A_\varepsilon \not = \emptyset, then

\Pi(A_\varepsilon \,|\, \{x_i\}_{i=1}^m) \rightarrow 1

almost surely as m \rightarrow \infty.

Remark on the asymptotics of the likelihood ratio and the K.-L. divergence

The problem.

Let f, g, h be three densities and suppose that, x_i \sim h, i \in \mathbb{N}, independently. What happens to the likelihood ratio

\prod_{i=1}^n \frac{f(x_i)}{g(x_i)}

as n\rightarrow \infty?

Clearly, it depends. If h = g \not = f, then

\prod_{i=1}^n \frac{f(x_i)}{g(x_i)} \rightarrow 0

almost surely at an exponential rate. More generally, if h is closer to g than to f, in some sense, we’d expect that \prod_{i=1}^n \frac{f(x_i)}{g(x_i)} \rightarrow 0. Such a measure of “closeness” of “divergence” between probability distributions is given by the Kullback-Leibler divergence

D_{KL}(f, g) = \int f \log\frac{f}{g}.

It can be verified that D_{KL}(f,g) \geq 0 with equality if and only if f=g, and that

D_{KL}(h,g) < D_{KL}(h,f) \Longrightarrow \prod_{i=1}^n \frac{f(x_i)}{g(x_i)} \rightarrow 0 \qquad (1)

almost surely at an exponential rate. Thus the K.L.-divergence can be used to solve our problem.

Better measures of divergence?

There are other measures of divergence that can determine the asymptotic behavior of the likelihood ratio as in (1) (e.g. the discrete distance). However, in this note, I give conditions under which the K.-L. divergence is, up to topological equivalence, the “best” measure of divergence.

This is of interest in Bayesian nonparametrics. The hypothesis that a density f_0 is in the Kullback-Leibler support of a prior \Pi on a density space \mathbb{F} is used to ensure that

\int_{\mathbb{F}} \prod_{i=1}^n \frac{f(x_i)}{f_0(x_i)} \Pi(df)

does not converge to 0 exponentially fast as n\rightarrow \infty and x_i \sim f_0. Under the conditions we specify, our remark implies that the hypothesis that “f_0 is in the Kullback-Leibler support of \Pi” may not be replaced by a weaker one.

Statement of the remark

Some notations. Let \mathcal{M} the space of all probability measures on some measurable space \mathbb{M}. If \mu, \nu \in \mathcal{M}, then both are absolutely continuous with respect to \tau = \mu + \nu and possess densities f, g such that d\mu = fd\tau and d\nu gd\tau. We denote by d\mu /d\nu the ratio of densities f / g, which in fact does not depend on the choice of dominating measure \tau. The likelihood ratio \prod_{i=1}^n \frac{f(x_i)}{g(x_i)} is abbreviated to \frac{d\mu}{d\nu}(X), depending implicitely on n. The Kullback-Leibler divergence between \mu and \nu is

D_{KL}(\mu, \nu) = \int \log\frac{d\mu}{d\nu} d\mu.

We let D be any other function \mathcal{M}\times \mathcal{M} \rightarrow [0,\infty] such that D(\mu, \nu) = 0 iff \mu = \nu and such that if D(\lambda, \nu) < D(\lambda, \mu), then there exists a \varepsilon > 0 with

e^{\varepsilon n} \frac{d\mu}{d\nu}(X) = e^{\varepsilon n} \prod_{i=1}^n \frac{d\mu}{d\nu}(x_i) \rightarrow 0

almost surely as n \rightarrow \infty. The topology on a subset \mathcal{F} \subset \mathcal{M} induced by such a function D is the topology induced by the sets

\{\nu \,|\, D(\mu, \nu) < \varepsilon\},\quad \mu \in \mathcal{F}, \, \varepsilon > 0.

The remark below shows that any exponential rate of convergence of the likelihood ratio is picked up by the KL divergence. It is rather obvious (albeit a bit technical), but I thought it was worth writing it up properly.

Let x_i \sim \mu, i \in \mathbb{N}, independently, and let \frac{d\nu}{d\mu}(X) = \prod_{i=1}^n \frac{d\nu}{d\mu}(x_i).

  1. We have that D_{KL}(\mu, \nu) < \infty if and only if \frac{d\nu}{d\mu}(X) does not converge more than exponentially fast to 0 (i.e. there exists \varepsilon > 0 such that e^{\varepsilon n}\frac{d\nu}{d\mu} (X) \rightarrow \infty).
  2. If \mathcal{F} \subset \mathcal{M} is such that D_{KL}(\mu, \nu) < \infty for all \mu, \nu \in \mathcal{F}, then

D(\lambda, \mu) < D(\lambda, \nu) \Longrightarrow D_{KL}(\lambda, \mu) < D_{KL}(\lambda, \nu)

and the topology on \mathcal{F} induced by D_{KL} is weaker than the topology of any other function D defined as above.

Proof of 1.
Suppose that D_{KL}(\mu, \nu) < \infty. Then, since by the strong law of large numbers D_{KL}(\mu, \nu) + \varepsilon - \frac{1}{n}\sum_{i=1}^n \log\frac{d\mu}{d\nu}(x_i) \rightarrow \varepsilon > 0, we find that

\log\left(e^{n (D_{KL}(\mu, \nu) + \varepsilon)} \frac{d\nu}{d\mu}(X)\right) = n (D_{KL}(\mu, \nu) + \varepsilon - \frac{1}{n}\sum_{i=1}^n \log\frac{d\mu}{d\nu}(x_i)) \rightarrow \infty

for all \varepsilon > 0.

If D_{KL}(\mu, \nu) = \infty, then for all \varepsilon > 0 we have

\log \left(e^{n \varepsilon}\frac{d\nu}{d\mu}(X)\right) = n\left(\varepsilon - \frac{1}{n}\sum_{i=1}^n \log\frac{d\mu(x_i)}{d\nu(x_i)}\right) \rightarrow -\infty

since \frac{1}{n}\sum_{i=1}^n \log\frac{d\mu(x_i)}{d\nu(x_i)} \rightarrow \infty. \Box

Proof of 2.
Suppose that there exists \lambda, \mu, \nu \in \mathcal{F} such that D(\lambda, \mu) < D(\lambda, \nu) but D_{KL}(\lambda, \mu) = D_{KL}(\lambda, \nu). Then, there is a \varepsilon > 0 such that

e^{2\varepsilon n} \frac{d\nu}{d\mu}(X)  = e^{n(D_{KL}(\lambda, \nu) + \varepsilon )} \frac{d\nu}{d\lambda}(X) /\left( e^{n(d_{KL}(\lambda, \mu) - \varepsilon)} \frac{d\mu}{d\lambda}(X) \right) \rightarrow 0.

But e^{n(D_{KL}(\lambda, \nu) + \varepsilon )} \frac{d\nu}{d\lambda}(X) \rightarrow \infty and e^{n(d_{KL}(\lambda, \mu) - \varepsilon)} \frac{d\mu}{d\lambda}(X) \rightarrow 0, which yields the contradiction.

Since D(\lambda, \mu) =0 iff \lambda = \mu, this implies that the topology on \mathcal{F} induced by D_{KL} is weaker than the one induced by D. \Box

The choice of prior in bayesian nonparametrics – Introduction

In preparation for the 11th Bayesian nonparametrics conference, I’m writing (and rewriting) notes on the background of our research (i.e. some of the general theory of bayesian nonparametrics). There are some good books on the subject (such as Bayesian Nonparametrics (Ghosh and Ramamoorthi, 2003)), but I wanted a more introductory focus and to present Choi and Ramamoorthi’s very clear point of view on posterior consistency (Remarks on the consistency of posterior distributions, 2008).

1. Introduction

Let \mathbb{X} be a complete and separable metric space and let \mathcal{M} be the space of all probability measures on \mathbb{X}. Some unknown distribution P_0\in \mathcal{M} is generating observable data \mathcal{D}_n = (X_1, X_2, \dots, X_n) \in \mathbb{X}^n, where each X_i is independently drawn from P_0. The problem is to learn about P_0 using only \mathcal{D}_n and prior knowledge.

Example (Discovery probabilities).
A cryptographer observes words, following some distribution P_0, in an unknown countable language \mathcal{L}. What are the P_0-probabilities of the words observed thus far? What is the probability that the next word to be observed has never been observed before?

1.1 Learning and uncertainty

We need an employable definition of learning. As a first approximation, we can consider learning to be the reduction of uncertainty about what is P_0. This requires a quantification of how uncertain we are to begin with. Then, hopefully, as data is gathered out uncertainty decreases and we are able to pinpoint P_0.

This is the core of Bayesian learning, alghough our definition is not yet entirely satisfactory. There are some difficulties with this idea of quantifying uncertainty, at least when using information-theoric concepts. The solution we adopt here is the use of probabilities to quantify uncertain knowledge (bayesians would also talk of subjective probabilities quantifying rational belief). For example, you may know that a coin flip is likely to be fair, although it is not impossible the two sides of the coin are both the same. This is uncertain knowledge about the distribution of heads and tails in the coin flips, and you could assign probabilities to the different possibilities.

More formally, prior uncertain knowledge about what is P_0 is quantified by a probability measure \Pi on \mathcal{M}. For any A \subset \mathcal{M}, \Pi(A) is the the prior probability that “P_0 \in A“. Then, given data \mathcal{D}_n, prior probabilities are adjusted to posterior probabilities: \Pi becomes \Pi_n, the conditional distribution of \Pi given \mathcal{D}_n. The celebrated Bayes’ theorem provides a formula to calculate \Pi_n from \Pi and \mathcal{D}_n. Thus we have an operational definition of learning in our statistical framework.

Learning is rationally adjusting uncertain knowledge in the light of new information.

For explanations as to why probabilities are well suited to the representation of uncertain knowledge, I refer the reader to Pearl (Probabilistic Reasoning in Intelligent Systems: Networks of Plausible Inference, 1988). We will also see that the operation of updating the prior to posterior posterior probabilities does work as intended.

1.2 The choice of prior

Specifying prior probabilities, that is quantifying prior uncertain knowledge, is not a simple task. It is especially difficult when uncertainty is over the non-negligeable part \mathcal{M} of an infinite dimensional vector space. Fortunately, “probability is not about numbers, it is about the structure of reasoning”, as Glenn Shafer puts it (cited in Pearl, 1988, p.15). The exact numbers given to the events “P_0 \in A” are not of foremost importance; what matters is how probabilities are more qualitatively put together, and how this relates to the learning process.

Properties of prior distributions, opening them to scrutiny, criticism and discussion, must be identified and related to what happens as more and more data is gathered.

Part 2.

Constrained semiparametric modelling (for directional statistics)



Angular data arises in many scientific fields, such as in experimental biology for the study of animal orientation, and in bioinformatics in relation to the protein structure prediction problem.



The statistical analysis of this data requires adapted tools such as 2\pi-periodic density models. Fernandez-Duran (Biometrics, 60(2), 2004) proposed non-negative trigonometric sums (i.e. non-negative trigonometric polynomials) as a flexible family of circular distributions. However, the coefficients of trigonometric polynomials expressed in the standard basis 1, \cos(x), \sin(x), \dots are difficult to interpret and we do not see how an informative prior could be specified through this parametrization. Moreover, the use of this basis was criticized by Ferreira et al. (Bayesian Analysis, 3(2), 2008) as resulting in a “wigly approximation, unlikely to be useful in most real applications”.

Trigonometric density basis

Here, we suggest the use of a density basis of the trigonometric polynomials and argue it is well suited to statistical applications. In particular, coefficients of trigonometric densities expressed in this basis possess an intuitive geometric interpretation. Furthermore, we show how “wiggliness” can be precisely controlled using this basis and how another geometric constraint, periodic unimodality, can be enforced [first proposition on the poster]. To ensure that nothing is lost by using this basis, we also show that the whole model consists of precisely all positive trigonometric densities, together with the basis functions [first theorem on the poster].

Prior specification

Priors can be specified on the coefficients of mixtures in our basis and on the degree of the trigonometric polynomials to be used. Through the interpretability of the coefficients and the shape-preserving properties of the basis, different types of prior knowledge may be incorporated. Together with an approximate understanding of mass allocation, these include:

  • periodic unimodality;
  • bounds on total variation; and
  • knowledge of the marginal distributions (in the multivariate case).

The priors obtained this way are part of a well-studied family called sieve priors, including the well-known Bernstein-Dirichlet prior, and are finite mixtures with an unknown number of components. Most results and interpretations about the Bernstein-Dirichlet prior (see Petrone & Wasserman (J. R. Stat. Soc. B., 64(1),  2002), Kruijer and Van der Vaart (J. Stat. Plan. Inference, 138(7), 2008), McVinish et al. (Scand. J. Statist., 36(2), 2009) can carry over to the priors we consider, but we dot not discuss them further.

Approximation-theoric framework

Our density models arise as the image of “shape-perserving” linear approximation operators. This approximation-theoric relationship is used to obtain a notably large prior Kullback-Leibler support and ensures strong posterior consistency at all bounded (not necessarily continuous) density. The result partly relies on known properties of sieve priors, as well as general consistency results (Walker (Ann. Statist., 32(5), 2004)), but extends known result by removing an usual continuity hypothesis on the densities at which consistency is achieved (see Wu & Ghosal (‎Electron. J. Stat., 2, 2008), Petrone & Veronese (Statistica Sinica, 20, 2010)). For contraction rates, higher order smoothness conditions are usually required (see Shen & Ghosal (Scand. J. Statist., 42(4), 2015)).

For example, consider the prior induced by the random density

T_n \mathcal{D} := \sum_i \mathcal{D}(R_{i,n}) C_{i,n},\qquad (1)

where \mathcal{D} is a Dirichlet process, n is distributed on \mathbb{N} and R_{i,n} is a partition of the circle. It has the strong posterior consistency at all bounded density provided that the associated operator

T_n : f \mapsto \sum_i C_{i,n} \int_{R_{i,n}} f

is such that \|T_n f - f\|_\infty \rightarrow 0 for all continuous f.

More generally, let \mathbb{F} be a set of bounded densities on some compact metric space \mathbb{M}, let T_n : L^1(\mathbb{M}) \rightarrow L^1(\mathbb{M}), n \in \mathbb{N}, be a sequence of operators that are:

  • shape preserving: T_n maps densities to densities and T_n(\mathbb{F}) \subset \mathbb{F}; and
  • approximating: \|T_n f - f\|_\infty \rightarrow 0 for all continuous f;

and finally let \Pi_n be priors on T_n(\mathbb{F}) with full support. A sieve prior on \mathbb{F} is defined by

\Pi : A \mapsto \sum_n \rho(n) \Pi_n(A \cap T_n(\mathbb{F})).

If 0 < \rho(n) < Ce^{-c d_n} for some increasing sequence d_n bounding the dimensions of T_n (\mathbb{F}), then the posterior distribution of \Pi is strongly consistent at each density of \mathbb{F}.

The approximation theory literature is rich in such operators. The theorem shows that they provide strongly consistent priors on arbitrary density spaces simply given priors \Pi_n on T_n(\mathbb{F}).

Basic density estimation:


A thousand samples (grey histogram) were drawn from the density in orange. The prior is defined by (1) with the Dirichlet process centered on the uniform density and with a precision parameter of 2. The degree n is distributed as a \text{Poiss}(15). The blue line is the posterior mean, the dark blue shaded region is a 50% pointwise credible region around the median, and the light blue shaded region is a 90% credible region.

Comment on The Sample Size Required in Importance Sampling

I summarize and comment part of The Sample Size Required in Importance Sampling (Chatterjee and Diaconis, 2015). One innovative idea is to bound the mean estimation error in terms of the tail behavior of d\mu/d\lambda, where \mu and \lambda are the importance sampling target and proposal distributions, respectively.

The problem is to evaluate

I = I(f) = \int f d\mu,

where \mu is a probability measure on a space \mathbb{M} and where f: \mathbb{M} \rightarrow \mathbb{R} is measurable. The Monte-Carlo estimate of I is

\frac{1}{n}\sum_{i=1}^n f(x_i), \qquad x_i \sim \mu.

When it is too difficult to sample \mu, for instance, other estimates can be obtained. Suppose that \mu is absolutely continuous with respect to another probability measure \lambda, and that the density of \mu with respect to \lambda is given by \rho. Another unbiaised estimate of I is then

I_n(f) = \frac{1}{n}\sum_{i=1}^n f(y_i)\rho(y_i),\qquad y_i \sim \lambda.

This is the general framework of importance sampling, with the Monte-Carlo estimate recovered by taking \lambda = \mu. An important question is the following.

How large should n be for I_n(f) to be close to I(f)?

An answer is given, under certain conditions, by Chatterjee and Diaconis (2015). Their main result can be interpreted as follows. If X \sim \mu and if \log \rho(X) is concentrated around its expected value L=\text{E}[\log \rho(X)], then a sample size of approximately n = e^{L} is both necessary and sufficient for I_n to be close to I. The exact sample size needed depends on \|f\|_{L^2(\mu)} and on the tail behavior of \log\rho(X). I state below their theorem with a small modification.

Theorem 1. (Chatterjee and Diaconis, 2015)
As above, let X \sim \mu. For any a > 0 and n \in \mathbb{N},

\mathbb{E} |I_n(f) - I(f)| \le \|f\|_{L^2(\mu)}\left( \sqrt{a/n} + 2\sqrt{\mathbb{P} (\rho(X) > a)} \right).

Conversely, for any \delta \in (0,1) and b > 0,

\mathbb{P}(1 - I_n(1) \le \delta) \le \frac{n}{b} + \frac{\mathbb{P}(\rho(X) \le b)}{1-\delta}.

Remark 1.
Suppose \|f\|_{L^2(\mu)} \le 1 and that \log\rho(X) is concentrated around L = \mathbb{E} \log\rho(X), meaning that for some t > 0 we have that \mathbb{P}(\log \rho(X) > L+t/2) and \mathbb{P}(\log\rho(X) < L-t/2) are both less than an arbitrary \varepsilon > 0. Then, taking n \geq e^{L+t} we find

\mathbb{E} |I_n(f) - I| \le e^{-t/4} + 2\varepsilon.

However, if n \leq e^{L-t} , we obtain

\mathbb{P}\left(1 - I_n(1) \le \tfrac{1}{2}\right) \le e^{-t/2} + 2 \varepsilon.

meaning that there can be a  high probability that I(1) and I_n(1) are not close.

Remark 2.
Let \lambda = \mu, so that \rho = 1. In that case, \log\rho(X) only takes its expected value 0. The theorem yields

\mathbb{E} |I_n(f) - I(f)| \le \frac{\|f\|_{L^2(\mu)}}{\sqrt{n}}

and no useful bound on \mathbb{P}(1-I_n(1) \le \delta).

For the theorem to yield a sharp cutoff, it is necessary that L = \mathbb{E} \log\rho(X) be relatively large and that \log\rho(X) be highly concentrated around L. The first condition is not aimed at in the practice of importance sampling. This difficulty contrasts with the broad claim that “a sample of size approximately e^{L} is necessary and sufficient for accurate estimation by importance sampling”. The result in conceptually interesting, but I’m not convinced that a sharp cutoff is common.


I consider their example 1.4. Here \lambda is the exponential distribution of mean 1, \mu is the exponential distribution of mean 2,\rho(x) = e^{x/2}/2 and f(x) = x. Thus I(f) = 2. We have L = \mathbb{E}\log\rho(X) = 1-\log(2) \approxeq 0.3, meaning that the theorem yields no useful cutoff. Furthermore, {}\mathbb{P}(\rho(X) > a) = \tfrac{1}{2a} and \|f\|_{L^2(\mu)} = 2. Optimizing the bound given by the theorem yields

\mathbb{E}|I_n(f)-2| \le \frac{4\sqrt{2}}{(2n)^{1/4}}.

The figure below shows 100 trajectories of I_k(f). The shaded area bounds the expected error.


This next figure shows 100 trajectories for the Monte-Carlo estimate of 2 = \int x d\mu, taking \lambda = \mu and \rho = 1. Here the theorem yields

\mathbb{E}|I_n(f)-2| \le \frac{2}{\sqrt{n}}.



Chatterjee, S. and Diaconis, P. The Sample Size Required in Importance Sampling.

Linear approximation operators and statistical models

We discuss the approximation properties of sequences of linear operators T_n mapping densities to densities. We give conditions for their convergence, explicit their general form, obtain rates of convergences and generalise the index parameter to obtain nets \{T_n\}_{n \in N}.

Notations. Let (\mathbb{M}, d) be a compact metric space, equipped with a finite measure \mu defined on its Borel \sigma-algebra, and denote by \mathcal{F} \subset L^1 the set of all essentially bounded probability densities on \mathbb{M}. The set \mathcal{F} is then a complete separable metric space under the total variation distance proportional to || f-g ||_1 = \int |f-g| d\mu.

In bayesian statistics, it is of interest to specify a probability measure P on \mathcal{F}, representing uncertainty about which distribution of \mathcal{F} is generating independent observations x_i \in \mathbb{M}. The problem is that \mathcal{F} is usually rather big: by Baire’s category theorem, if \mathbb{M} is not a finite set of points, then \mathcal{F} cannot be written as a countable union of finite dimensional subspaces. To help in prior elicitation, that is to help a statistician specify P, we may decompose \mathcal{F} in simpler parts.

Here, I discuss how to obtain a sequence of approximating finite dimensional sieves \mathcal{S}_n \subset \mathcal{F}, such that \cup_n \mathcal{S}_n is dense in \mathcal{F}. A prior P on \mathcal{F} may then be specified as the countable mixture

P = \sum _{n \geq 1} \alpha_n P_{\mathcal{S}_n}, \quad \alpha_n \geq 0,\, \sum_n \alpha_n = 1,

where P_{\mathcal{S}_n} is a prior on \mathcal{S}_n for all n.

Let me emphasize that the following ideas are elementary.  Some may be found, with more or less generality, in analysis and approximation theory textbooks. It is, however, interesting to recollect the facts relevant in statistical applications.

1. The basics

The finite dimensional sieves \mathcal{S}_n take the form

\mathcal{S}_n =  \left\{ \sum_{i=0}^{m_n} c_i \phi_{i,n} \right\}, \quad m_n \in \mathbb{N}

where the \phi_{i,n} are densities and the coefficients c_i range through some set which we assume contains the simplex \Delta_n = \left\{ (c_i) : \sum c_i = 1,\, c_i \geq 0 \right\}.

The following lemma gives sufficient conditions for \cup_n \mathcal{S}_n to be dense in \mathcal{F}, with the total variation distance.

Continue reading

The asymptotic behaviour of posterior distributions – 1

I describe key ideas of the theory of posterior distribution asymptotics and work out small examples.

1. Introduction

Consider the problem of learning about the probability distribution that a stochastic mechanism is following, based on independent observations. You may quantify your uncertainty about what distribution the mechanism is following through a prior P defined on the space of all possible distributions, and then obtain the conditional distribution of P given the observations to correspondingly adjust your uncertainty. In the simplest case, the prior is concentrated on a space \mathcal{M} of distributions dominated by a common measure \lambda. This means that each probability measure in \mu \in \mathcal{M} is such that there exists f with \mu(E) = \int_E f \,d\lambda for all measurable E. We may thus identify \mathcal{M} with a space \mathbb{F} of probability densities. Given independent observations X = (X_1, \dots, X_n), the conditional distribution of P given X is then

P(A \mid X) = \frac{\int_A f(X) P(df)}{\int_{\mathbb{F}} f(X) P(df)}, \quad A \subset \mathbb{F},

where f(X) = \Pi_{i=1}^nf(X_i) and A is measurable. This conditional distribution is called the posterior distribution of P, and is understood as a Lebesgue integral relative to the measure P defined on \mathbb{F} \simeq \mathcal{M}.

The procedure of conditionning P on X is bayesian learning. It is expected that as more and more data is gathered, that the posterior distribution will converge, in a suitable way, to a point mass located at the true distribution that the data is following. This is the asymptotic behavior we study.

1.1 Notations

The elements of measure theory and the language of mathematical statistics I use are standard and very nicely reviewed in Halmos (1949). To keep notations light, I avoid making explicit references to measure spaces. All sets and maps considered are measurable.

2. The relative entropy of probability measures

Let \mu_1 and \mu_2 be two probability measures on a metric space \mathbb{M}, and let \lambda be a common \sigma-finite dominating measure such as \mu_1 + \mu_2. That is, \mu_i(E) = 0 whenever E \subset \mathbb{M} is of \lambda-measure 0, and by the Radon-Nikodym theorem this is equivalent to the fact that there exists unique densities f_1, f_2 with d\mu_i = f_i d\lambda. The relative entropy of \mu_1 and \mu_2, also known as the Kullback-Leibler divergence, is defined as

K(\mu_1, \mu_2) = \int \log \frac{f_1}{f_2} d\mu_1\,.

The following inequality will be of much use.

Lemma 1 (Kullback and Leibler (1951)). We have K(\mu_1, \mu_2) \geq 0, with equality if and only if \mu_1 = \mu_2.

Proof: We may assume that \mu_2 is dominated by \mu_1, as otherwise K(\mu_1, \mu_2) = \infty and the lemma holds. Now, let \phi (t) = t\log t, g = f_1/f_2 and write {}K(\mu_1, \mu_2) = \int \phi \circ g d\mu_2. Since \phi (t) = (t-1) +(t-1)^2 / h(t) for some h(t) > 0, we find

K(\mu_1, \mu_2) = \int (g-1) d\mu_2 + \int \frac{(g-1)^2}{h} d\mu_2 = \int \frac{(g-1)^2}{h\circ g} d\mu_2 \geq 0,

with equality if and only if f_1/f_2 = 1, \mu_1-almost everywhere. QED.

We may already apply the lemma to problems of statistical inference.

2.1 Discriminating between point hypotheses

Alice observes X = (X_1, \dots, X_n), X_i \sim^{ind}\mu_*, \mu_* \in \{\mu_1, \mu_2\}, and considers the hypotheses H_i = \{\mu_i\}, i=1,2, representing “\mu_* = \mu_i“. A prior distribution on H_1 \cup H_2 takes the form

P = \alpha \delta_1 + (1-\alpha)\delta_2, \quad 0 < \alpha < 1,

where \delta_i(A) is 1 when \mu_i \in A and is 0 otherwise. Given a sample X_1 \sim \mu_1, the weight of evidence for H_1 versus H_2 (Good, 1985), also known as “the information in X_1 for discrimination between H_1 and H_2” (Kullback, 1951), is defined as

W(X_1) = \log\frac{f_1(X_1)}{f_2(X_1)}.

This quantity is additive for independent sample: if X = (X_1, \dots, X_n), X_i \sim^{ind} \mu_*, then

W(X) = \sum_{i=1}^n W(X_i);

and the posterior log-odds are given by
\log\frac{P(H_1 \mid X)}{P(H_2 \mid X)} = W(X) + \log\frac{P(H_1)}{P(H_2)}.
Thus K(\mu_1, \mu_2) is the expected weight of evidence for H_1 against H_2 brought by an unique sample X_1 \sim \mu_1 and is strictly positive whenever \mu_1 \not = \mu_2. By the additivity of W, Alice should expect that as more and more data is gathered, the weight of evidence grows to infinity. In fact, this happens \mu_1-almost surely.

Proposition 2. Almost surely as the number of observations grows, W(X) \rightarrow \infty and P(H_1 | X)\rightarrow 1.

Proof: By the law of large numbers (the case K(\mu_1, \mu_2) = \infty is easily treated separatly), we have

\frac{1}{n}\sum_{i=1}^n \log\frac{f_1(X_i)}{f_2(X_i)}\rightarrow K(\mu_1, \mu_2) > 0, \quad \mu_1\text{-a.s.}.


W(X) = n \left( \frac{1}{n}\sum_{i=1}^n \log\frac{f_1(X_i)}{f_2(X_i)} \right) \rightarrow \infty \quad \mu_1\text{-a.s.}

and P(H_1\mid X) \rightarrow 1 \mu_1-almost surely. QED.

2.2 Finite mixture prior under misspecification

Alice wants to learn about a fixed unknown distribution \mu_* through data X =(X_1, \dots, X_n), where X_i \sim^{ind.} \mu_*. She models what \mu_* may be as one of the distribution in \mathcal{M} = \{\mu_1, \dots, \mu_k\}, and quantifies her uncertainty through a prior P on \mathcal{M}. We may assume that K(\mu_*, \mu_i) < \infty for some i, as otherwise she will eventually observe data that is impossible under \mathcal{M} and adjust her model. (Indeed, K(\mu_*, \mu_i)=\infty implies that there exists a \mu_* non-negligible set E_i such that \mu_i(E_i) = 0 < \mu_*(E_i). Alice will \mu_*-almost surely observe X_m \in E_i and conclude that \mu_* \not = \mu_k.) If \mu_* \not\in \mathcal{M}, she may not realise that her model is wrong, but the following proposition ensures that the posterior distribution will concentrate on the \mu_i‘s that are closest to \mu_*.

Proposition 3. Let A = \{\mu_i \in \mathcal{M} \mid K(\mu_*, \mu_i) = \min_j K(\mu_*, \mu_j)\}. Almost surely as the number of observations grows, {} P(A | X) \rightarrow 1.

Proof: The prior takes the form {}P = \sum_{i=1}^{k} \alpha_i \delta_i, {}\alpha_i > 0, where {}\delta_i is a point mass at {}\mu_i. The model {}\mathcal{M} is dominated by a {}\sigma-finite measure {}\lambda, such as {}\sum \mu_i, so that the posterior distribution is

P(A\mid X) = \frac{\sum_{\mu_i \in A} f_i(X) \alpha_i }{\sum_{\mu_i \in \mathcal{M}} f_i(X) \alpha_i}, \quad d\mu_i = f_i d\lambda.

Because {}K(\mu_\star, \mu_i) < \infty for some {}i, {}\mu_\star is also absolutely continuous with respect to {}\lambda and we let {}d\mu_\star = f_\star d\lambda. Now, let {}\varepsilon = \min_j K(\mu_\star, \mu_j) and {}\delta = \min_j \{K(\mu_\star, \mu_j \mid \mu_j \in A^c)\} > \varepsilon. Write

\log \frac{P(A | X)}{P(A^c | X)} \geq \log \frac{f_A(X)}{f_{A^c}(X)} + \log \frac{\sum_{\mu_i \in A} \alpha_i}{ \sum_{\mu_i \in A^c} \alpha_i},

where {}f_A = \arg \min_{f_i : \mu_i \in A} f_i(X) and {}f_{A^c} = \arg \max_{f_i : \mu_i \in A^c} f_i(X) both depend on {}n. Using the law of large numbers, we find

\liminf_{n \rightarrow \infty} \frac{1}{n} \sum_{i=1}^n \log\frac{f_\star(X_i)}{f_{A^c}(X_i)} \geq \delta > \varepsilon = \lim_{n \rightarrow \infty} \frac{1}{n}\sum_{i=1}^n \log\frac{f_\star(X_i)}{f_A (X_i)}\quad \mu_\star\text{-a.s.},

and therefore

{}\log \frac{f_A(X)}{f_{A^c}(X)} = n\left( \frac{1}{n} \sum_{i=1}^n \log\frac{f_\star(X_i)}{f_{A^c}(X_i)} - \frac{1}{n}\sum_{i=1}^n \log\frac{f_\star(X_i)}{f_A (X_i)} \right) \rightarrow \infty\quad \mu_*\text{-a.s.}.

This implies {}P(A \mid X) \rightarrow 1, {}\mu_\star-almost surely. QED.

2.3 Properties of the relative entropy

The following properties justify the interpretation of the relative entropy K(\mu_1, \mu_2) as an expected weight of evidence.

Lemma 4. The relative entropy {}K(\mu_1, \mu_2) = \int f_1 \log\frac{f_1}{f_2} d\lambda, {}d\mu_i = f_id\lambda, does not depend on the choice of dominating measure {}\lambda.

Let {}T: \mathbb{M} \rightarrow \mathbb{M}' be a mapping onto a space {}\mathbb{M}'. If {}X \sim \mu, then {}T(X) \sim \mu T^{-1}, where {}\mu T^{-1} is the probability measure on {}\mathbb{M}' defined by {}\mu T^{-1}(F) = \mu (T^{-1}(F)). The following proposition, a particular case of (Kullback, 1951, theorem 4.1), states that transforming the data through {}T cannot increase the expected weight of evidence.

Proposition 5 (see Kullback and Leibler (1951)). For any two probability measures {}\mu_1, \mu_2 on {}\mathbb{M}, we have

{}K(\mu_1 T^{-1}, \mu_2 T^{-1}) \le K(\mu_1, \mu_2).

Proof:  Let {}\lambda be a dominating measure for {}\{\mu_1, \mu_2\}, {}d\mu_i = f_i d\lambda. Therefore, {}\mu_i T^{-1} is dominated by {}\lambda T^{-1}, and we may write {}d\mu_i T^{-1} = g_i d\lambda T^{-1}, {}i=1,2. The measures on the two spaces are related by the formula

{} \int_{T^{-1}(F)} h \circ T \, d\alpha = \int_{F} h \,d\alpha T^{-1},

where {}\alpha may be one of {}\mu_1, {}\mu_2 and {}\lambda and {}h is any measurable function (Halmos, 1949, lemma 3). Therefore,

{}K(\mu_1, \mu_2) - K(\mu_1 T^{-1}, \mu_2 T^{-1}) = \int_{\mathbb{M}} \log\frac{f_1}{f_2} d\mu_1 - \int_{\mathbb{M}'} \log \frac{g_1}{g_2} d\mu_1T^{-1} \\ = \int \log\frac{f_1}{f_2} d\mu_1 - \int \log\frac{g_1 \circ T}{g_2 \circ T} d\mu_1 = (\star).

By letting {}h = \frac{f_1 g_2 \circ T}{f_2 g_1 \circ T} we find

{} (\star) = \int h \log h \, \frac{g_1\circ T}{g_2 \circ T} d\mu_2 = \int h \log h \,d\gamma,

where {}\gamma is such that {}d\gamma = \frac{g_1\circ T}{g_2 \circ T} d\mu_2. It is a probability measure since {}\int \frac{g_1\circ T}{g_2 \circ T} d\mu_2 = 1.

By convexity of {}\phi : t \mapsto t \log t, we find

{} \int h \log h \,d\gamma \geq \phi\left( \int h \,d\gamma \right) = \phi(1) = 0

and this finishes the proof.


Good, I. (1985). Weight of evidence: A brief survey. In A. S. D. L. J.M. Bernardo, M.H. DeGroot (Ed.), Bayesian Statistics 2, pp. 249–270. North-Holland B.V.: Elsevier Science Publishers.

Halmos, P. R. and L. J. Savage (1949, 06). Application of the radon-nikodym theorem to the theory of sufficient statistics. Ann. Math. Statist. 20 (2), 225–241.

Kullback, S. and Leibler, R. A. (1951). On information and sufficiency. Ann. Math. Statist. 22(1), 79-86