## Bayesian conjugate gradients [open for discussion]

Posted in Books, pictures, Statistics, University life with tags , , , , , on June 25, 2019 by xi'an

When fishing for an illustration for this post on Google, I came upon this Bayesian methods for hackers cover, a book about which I have no clue whatsoever (!) but that mentions probabilistic programming. Which serves as a perfect (?!) introduction to the call for discussion in Bayesian Analysis of the incoming Bayesian conjugate gradient method by Jon Cockayne, Chris Oates (formerly Warwick), Ilse Ipsen and Mark Girolami (still partially Warwick!). Since indeed the paper is about probabilistic numerics à la Mark and co-authors. Surprisingly dealing with solving the deterministic equation Ax=b by Bayesian methods. The method produces a posterior distribution on the solution x⁰, given a fixed computing effort, which makes it pertain to the anytime algorithms. It also relates to an earlier 2015 paper by Christian Hennig where the posterior is on A⁻¹ rather than x⁰ (which is quite a surprising if valid approach to the problem!) The computing effort is translated here in computations of projections of random projections of Ax, which can be made compatible with conjugate gradient steps. Interestingly, the choice of the prior on x is quite important, including setting a low or high convergence rate…  Deadline is August 04!

## the end of the Series B’log…

Posted in Books, Statistics, University life with tags , , , , on September 22, 2017 by xi'an

Today is the last and final day of Series B’log as David Dunson, Piotr Fryzlewicz and myself have decided to stop the experiment, faute de combattants. (As we say in French.) The authors nicely contributed long abstracts of their papers, for which I am grateful, but with a single exception, no one came out with comments or criticisms, and the idea to turn some Series B papers into discussion papers does not seem to appeal, at least in this format. Maybe the concept will be rekindled in another form in the near future, but for now we let it lay down. So be it!

## beyond objectivity, subjectivity, and other ‘bjectivities

Posted in Statistics with tags , , , , , , , , , , , , , on April 12, 2017 by xi'an

Here is my discussion of Gelman and Hennig at the Royal Statistical Society, which I am about to deliver!

## objective and subjective RSS Read Paper next week

Posted in Books, pictures, Statistics, Travel, University life, Wines with tags , , , , , , , , , , , , , , on April 5, 2017 by xi'an

Andrew Gelman and Christian Hennig will give a Read Paper presentation next Wednesday, April 12, 5pm, at the Royal Statistical Society, London, on their paper “Beyond subjective and objective in statistics“. Which I hope to attend and else to write a discussion. Since the discussion (to published in Series A) is open to everyone, I strongly encourage ‘Og’s readers to take a look at the paper and the “radical” views therein to hopefully contribute to this discussion. Either as a written discussion or as comments on this very post.

## a response by Ly, Verhagen, and Wagenmakers

Posted in Statistics with tags , , , , , , , , on March 9, 2017 by xi'an

Following my demise [of the Bayes factor], Alexander Ly, Josine Verhagen, and Eric-Jan Wagenmakers wrote a very detailed response. Which I just saw the other day while in Banff. (If not in Schiphol, which would have been more appropriate!)

“In this rejoinder we argue that Robert’s (2016) alternative view on testing has more in common with Jeffreys’s Bayes factor than he suggests, as they share the same ‘‘shortcomings’’.”

Rather unsurprisingly (!), the authors agree with my position on the dangers to ignore decisional aspects when using the Bayes factor. A point of dissension is the resolution of the Jeffreys[-Lindley-Bartlett] paradox. One consequence derived by Alexander and co-authors is that priors should change between testing and estimating. Because the parameters have a different meaning under the null and under the alternative, a point I agree with in that these parameters are indexed by the model [index!]. But with which I disagree when arguing that the same parameter (e.g., a mean under model M¹) should have two priors when moving from testing to estimation. To state that the priors within the marginal likelihoods “are not designed to yield posteriors that are good for estimation” (p.45) amounts to wishful thinking. I also do not find a strong justification within the paper or the response about choosing an improper prior on the nuisance parameter, e.g. σ, with the same constant. Another a posteriori validation in my opinion. However, I agree with the conclusion that the Jeffreys paradox prohibits the use of an improper prior on the parameter being tested (or of the test itself). A second point made by the authors is that Jeffreys’ Bayes factor is information consistent, which is correct but does not solved my quandary with the lack of precise calibration of the object, namely that alternatives abound in a non-informative situation.

“…the work by Kamary et al. (2014) impressively introduces an alternative view on testing, an algorithmic resolution, and a theoretical justification.”

The second part of the comments is highly supportive of our mixture approach and I obviously appreciate very much this support! Especially if we ever manage to turn the paper into a discussion paper! The authors also draw a connection with Harold Jeffreys’ distinction between testing and estimation, based upon Laplace’s succession rule. Unbearably slow succession law. Which is well-taken if somewhat specious since this is a testing framework where a single observation can send the Bayes factor to zero or +∞. (I further enjoyed the connection of the Poisson-versus-Negative Binomial test with Jeffreys’ call for common parameters. And the supportive comments on our recent mixture reparameterisation paper with Kaniav Kamari and Kate Lee.) The other point that the Bayes factor is more sensitive to the choice of the prior (beware the tails!) can be viewed as a plus for mixture estimation, as acknowledged there. (The final paragraph about the faster convergence of the weight α is not strongly

## nonparametric Bayesian clay for robust decision bricks

Posted in Statistics with tags , , , , , , on January 30, 2017 by xi'an

Just received an email today that our discussion with Judith of Chris Holmes and James Watson’s paper was now published as Statistical Science 2016, Vol. 31, No. 4, 506-510… While it is almost identical to the arXiv version, it can be read on-line.

## a Bayesian criterion for singular models [discussion]

Posted in Books, Statistics, University life with tags , , , , , , , , , , , , , , , , on October 10, 2016 by xi'an

[Here is the discussion Judith Rousseau and I wrote about the paper by Mathias Drton and Martyn Plummer, a Bayesian criterion for singular models, which was discussed last week at the Royal Statistical Society. There is still time to send a written discussion! Note: This post was written using the latex2wp converter.]

It is a well-known fact that the BIC approximation of the marginal likelihood in a given irregular model ${\mathcal M_k}$ fails or may fail. The BIC approximation has the form

$\displaystyle BIC_k = \log p(\mathbf Y_n| \hat \pi_k, \mathcal M_k) - d_k \log n /2$

where ${d_k }$ corresponds on the number of parameters to be estimated in model ${\mathcal M_k}$. In irregular models the dimension ${d_k}$ typically does not provide a good measure of complexity for model ${\mathcal M_k}$, at least in the sense that it does not lead to an approximation of

$\displaystyle \log m(\mathbf Y_n |\mathcal M_k) = \log \left( \int_{\mathcal M_k} p(\mathbf Y_n| \pi_k, \mathcal M_k) dP(\pi_k|k )\right) \,.$

A way to understand the behaviour of ${\log m(\mathbf Y_n |\mathcal M_k) }$ is through the effective dimension

$\displaystyle \tilde d_k = -\lim_n \frac{ \log P( \{ KL(p(\mathbf Y_n| \pi_0, \mathcal M_k) , p(\mathbf Y_n| \pi_k, \mathcal M_k) ) \leq 1/n | k ) }{ \log n}$

when it exists, see for instance the discussions in Chambaz and Rousseau (2008) and Rousseau (2007). Watanabe (2009} provided a more precise formula, which is the starting point of the approach of Drton and Plummer:

$\displaystyle \log m(\mathbf Y_n |\mathcal M_k) = \log p(\mathbf Y_n| \hat \pi_k, \mathcal M_k) - \lambda_k(\pi_0) \log n + [m_k(\pi_0) - 1] \log \log n + O_p(1)$

where ${\pi_0}$ is the true parameter. The authors propose a clever algorithm to approximate of the marginal likelihood. Given the popularity of the BIC criterion for model choice, obtaining a relevant penalized likelihood when the models are singular is an important issue and we congratulate the authors for it. Indeed a major advantage of the BIC formula is that it is an off-the-shelf crierion which is implemented in many softwares, thus can be used easily by non statisticians. In the context of singular models, a more refined approach needs to be considered and although the algorithm proposed by the authors remains quite simple, it requires that the functions ${ \lambda_k(\pi)}$ and ${m_k(\pi)}$ need be known in advance, which so far limitates the number of problems that can be thus processed. In this regard their equation (3.2) is both puzzling and attractive. Attractive because it invokes nonparametric principles to estimate the underlying distribution; puzzling because why should we engage into deriving an approximation like (3.1) and call for Bayesian principles when (3.1) is at best an approximation. In this case why not just use a true marginal likelihood?

1. Why do we want to use a BIC type formula?

The BIC formula can be viewed from a purely frequentist perspective, as an example of penalised likelihood. The difficulty then stands into choosing the penalty and a common view on these approaches is to choose the smallest possible penalty that still leads to consistency of the model choice procedure, since it then enjoys better separation rates. In this case a ${\log \log n}$ penalty is sufficient, as proved in Gassiat et al. (2013). Now whether or not this is a desirable property is entirely debatable, and one might advocate that for a given sample size, if the data fits the smallest model (almost) equally well, then this model should be chosen. But unless one is specifying what equally well means, it does not add much to the debate. This also explains the popularity of the BIC formula (in regular models), since it approximates the marginal likelihood and thus benefits from the Bayesian justification of the measure of fit of a model for a given data set, often qualified of being a Bayesian Ockham’s razor. But then why should we not compute instead the marginal likelihood? Typical answers to this question that are in favour of BIC-type formula include: (1) BIC is supposingly easier to compute and (2) BIC does not call for a specification of the prior on the parameters within each model. Given that the latter is a difficult task and that the prior can be highly influential in non-regular models, this may sound like a good argument. However, it is only apparently so, since the only justification of BIC is purely asymptotic, namely, in such a regime the difficulties linked to the choice of the prior disappear. This is even more the case for the sBIC criterion, since it is only valid if the parameter space is compact. Then the impact of the prior becomes less of an issue as non informative priors can typically be used. With all due respect, the solution proposed by the authors, namely to use the posterior mean or the posterior mode to allow for non compact parameter spaces, does not seem to make sense in this regard since they depend on the prior. The same comments apply to the author’s discussion on Prior’s matter for sBIC. Indeed variations of the sBIC could be obtained by penalizing for bigger models via the prior on the weights, for instance as in Mengersen and Rousseau (2011) or by, considering repulsive priors as in Petralia et al. (20120, but then it becomes more meaningful to (again) directly compute the marginal likelihood. Remains (as an argument in its favour) the relative computational ease of use of sBIC, when compared with the marginal likelihood. This simplification is however achieved at the expense of requiring a deeper knowledge on the behaviour of the models and it therefore looses the off-the-shelf appeal of the BIC formula and the range of applications of the method, at least so far. Although the dependence of the approximation of ${\log m(\mathbf Y_n |\mathcal M_k)}$ on ${\mathcal M_j }$, \$latex {j \leq k} is strange, this does not seem crucial, since marginal likelihoods in themselves bring little information and they are only meaningful when compared to other marginal likelihoods. It becomes much more of an issue in the context of a large number of models.

2. Should we care so much about penalized or marginal likelihoods ?

Marginal or penalized likelihoods are exploratory tools in a statistical analysis, as one is trying to define a reasonable model to fit the data. An unpleasant feature of these tools is that they provide numbers which in themselves do not have much meaning and can only be used in comparison with others and without any notion of uncertainty attached to them. A somewhat richer approach of exploratory analysis is to interrogate the posterior distributions by either varying the priors or by varying the loss functions. The former has been proposed in van Havre et l. (2016) in mixture models using the prior tempering algorithm. The latter has been used for instance by Yau and Holmes (2013) for segmentation based on Hidden Markov models. Introducing a decision-analytic perspective in the construction of information criteria sounds to us like a reasonable requirement, especially when accounting for the current surge in studies of such aspects.

[Posted as arXiv:1610.02503]