Archive for Dirichlet process

mixture models with a prior on the number of components

Posted in Books, Statistics, University life with tags , , , , , , , on March 6, 2015 by xi'an

mixdir

“From a Bayesian perspective, perhaps the most natural approach is to treat the numberof components like any other unknown parameter and put a prior on it.”

Another mixture paper on arXiv! Indeed, Jeffrey Miller and Matthew Harrison recently arXived a paper on estimating the number of components in a mixture model, comparing the parametric with the non-parametric Dirichlet prior approaches. Since priors can be chosen towards agreement between those. This is an obviously interesting issue, as they are often opposed in modelling debates. The above graph shows a crystal clear agreement between finite component mixture modelling and Dirichlet process modelling. The same happens for classification.  However, Dirichlet process priors do not return an estimate of the number of components, which may be considered a drawback if one considers this is an identifiable quantity in a mixture model… But the paper stresses that the number of estimated clusters under the Dirichlet process modelling tends to be larger than the number of components in the finite case. Hence that the Dirichlet process mixture modelling is not consistent in that respect, producing parasite extra clusters…

In the parametric modelling, the authors assume the same scale is used in all Dirichlet priors, that is, for all values of k, the number of components. Which means an incoherence when marginalising from k to (k-p) components. Mild incoherence, in fact, as the parameters of the different models do not have to share the same priors. And, as shown by Proposition 3.3 in the paper, this does not prevent coherence in the marginal distribution of the latent variables. The authors also draw a comparison between the distribution of the partition in the finite mixture case and the Chinese restaurant process associated with the partition in the infinite case. A further analogy is that the finite case allows for a stick breaking representation. A noteworthy difference between both modellings is about the size of the partitions

\mathbb{P}(s_1,\ldots,s_k)\propto\prod_{j=1}^k s_j^{-\gamma}\quad\text{versus}\quad\mathbb{P}(s_1,\ldots,s_k)\propto\prod_{j=1}^k s_j^{-1}

in the finite (homogeneous partitions) and infinite (extreme partitions) cases.

An interesting entry into the connections between “regular” mixture modelling and Dirichlet mixture models. Maybe not ultimately surprising given the past studies by Peter Green and Sylvia Richardson of both approaches (1997 in Series B and 2001 in JASA).

improved approximate-Bayesian model-choice method for estimating shared evolutionary history [reply from the author]

Posted in Books, Statistics, University life with tags , , , , , , , , , , , , on June 3, 2014 by xi'an

[Here is a very kind and detailed reply from Jamie Oakes to the comments I made on his ABC paper a few days ago:]

First of all, many thanks for your thorough review of my pre-print! It is very helpful and much appreciated. I just wanted to comment on a few things you address in your post.

I am a little confused about how my replacement of continuous uniform probability distributions with gamma distributions for priors on several parameters introduces a potentially crippling number of hyperparameters. Both uniform and gamma distributions have two parameters. So, the new model only has one additional hyperparameter compared to the original msBayes model: the concentration parameter on the Dirichlet process prior on divergence models. Also, the new model offers a uniform prior over divergence models (though I don’t recommend it).

Your comment about there being no new ABC technique is 100% correct. The model is new, the ABC numerical machinery is not. Also, your intuition is correct, I do not use the divergence times to calculate summary statistics. I mention the divergence times in the description of the ABC algorithm with the hope of making it clear that the times are scaled (see Equation (12)) prior to the simulation of the data (from which the summary statistics are calculated). This scaling is simply to go from units proportional to time, to units that are proportional to the expected number of mutations. Clearly, my attempt at clarity only created unnecessary opacity. I’ll have to make some edits.

Regarding the reshuffling of the summary statistics calculated from different alignments of sequences, the statistics are not exchangeable. So, reshuffling them in a manner that is not conistent across all simulations and the observed data is not mathematically valid. Also, if elements are exchangeable, their order will not affect the likelihood (or the posterior, barring sampling error). Thus, if our goal is to approximate the likelihood, I would hope the reshuffling would also have little affect on the approximate posterior (otherwise my approximation is not so good?).

You are correct that my use of “bias” was not well defined in reference to the identity line of my plots of the estimated vs true probability of the one-divergence model. I think we can agree that, ideally (all assumptions are met), the estimated posterior probability of a model should estimate the probability that the model is correct. For large numbers of simulation
replicates, the proportion of the replicates for which the one-divergence model is true will approximate the probability that the one-divergence model is correct. Thus, if the method has the desirable (albeit “frequentist”) behavior such that the estimated posterior probability of the one-divergence model is an unbiased estimate of the probability that the one-divergence model is correct, the points should fall near the identity line. For example, let us say the method estimates a posterior probability of 0.90 for the one-divergence model for 1000 simulated datasets. If the method is accurately estimating the probability that the one-divergence model is the correct model, then the one-divergence model should be the true model for approximately 900 of the 1000 datasets. Any trend away from the identity line indicates the method is biased in the (frequentist) sense that it is not correctly estimating the probability that the one-divergence model is the correct model. I agree this measure of “bias” is frequentist in nature. However, it seems like a worthwhile goal for Bayesian model-choice methods to have good frequentist properties. If a method strongly deviates from the identity line, it is much more difficult to interpret the posterior probabilites that it estimates. Going back to my example of the posterior probability of 0.90 for 1000 replicates, I would be alarmed if the model was true in only 100 of the replicates.

My apologies if my citation of your PNAS paper seemed misleading. The citation was intended to be limited to the context of ABC methods that use summary statistics that are insufficient across the models under comparison (like msBayes and the method I present in the paper). I will definitely expand on this sentence to make this clearer in revisions. Thanks!

Lastly, my concluding remarks in the paper about full-likelihood methods in this domain are not as lofty as you might think. The likelihood function of the msBayes model is tractable, and, in fact, has already been derived and implemented via reversible-jump MCMC (albeit, not readily available yet). Also, there are plenty of examples of rich, Kingman-coalescent models implemented in full-likelihood Bayesian frameworks. Too many to list, but a lot of them are implemented in the BEAST software package. One noteworthy example is the work of Bryant et al. (2012, Molecular Biology and Evolution, 29(8), 1917–32) that analytically integrates over all gene trees for biallelic markers under the coalescent.

improved approximate-Bayesian model-choice method for estimating shared evolutionary history

Posted in Books, Statistics, University life with tags , , , , , , , , , , , on May 14, 2014 by xi'an

“An appealing approach would be a comparative, Bayesian model-choice method for inferring the probability of competing divergence histories while integrating over uncertainty in mutational and ancestral processes via models of nucleotide substitution and lineage coalescence.” (p.2)

Jamies Oaks arXived (a few months ago now) a rather extensive Monte-Carlo study on the impact of prior modelling on the model-choice performances of ABC model choice. (Of which I only became aware recently.) As in the earlier paper I commented on the Óg, the issue here has much more to do with prior assessment and calibration than with ABC implementation per se. For instance, the above quote recaps the whole point of conducting Bayesian model choice. (As missed by Templeton.)

“This causes divergence models with more divergence-time parameters to integrate over a much greater parameter space with low likelihood yet high prior density, resulting in small marginal likelihoods relative to models with fewer divergence-time parameters.” (p.2)

This second quote is essentially stressing the point with Occam’s razor argument. Which I deem [to be] a rather positive feature of Bayesian model choice. A reflection on the determination of the prior distribution, getting away from uniform priors, thus sounds most timely! The current paper takes place within a rather extensive exchange between Oak’s group and Hickerson’s group on what makes Bayesian model choice (and the associated software msBayes) pick or not the correct model. Oak and coauthors objected to the use of “narrow, empirically informed uniform priors”, arguing that this leads to a bias towards models with less parameters, a “statistical issue” in their words, while Hickerson et al. (2014) think this is due to msBayes way of selecting models and their parameters at random. However it refrains from reproducing earlier criticisms of or replies to Hickerson et al.

The current paper claims to have reached a satisfactory prior modelling with ¨improved robustness, accuracy, and power” (p.3).  If I understand correctly, the changes are in replacing a uniform distribution with a Gamma or a Dirichlet prior. Which means introducing a seriously large and potentially crippling number of hyperparameters into the picture. Having a lot of flexibility in the prior also means a lot of variability in the resulting inference… In other words, with more flexibility comes more responsibility, to paraphrase Voltaire.

“I have introduced a new approximate-Bayesian model choice method.” (p.21)

The ABC part is rather standard, except for the strange feature that the divergence times are used to construct summary statistics (p.10). Strange because these times are not observed for the actual data. So I must be missing something. (And I object to the above quote and to the title of the paper since there is no new ABC technique there, simply a different form of prior.)

“ABC methods in general are known to be biased for model choice.” (p.21)

I do not understand much the part about (reshuffling) introducing bias as detailed on p.11: every approximate method gives a “biased” answer in the sense this answer is not the true and proper posterior distribution. Using a different (re-ordered) vector of statistics provides a different ABC outcome,  hence a different approximate posterior, for which it seems truly impossible to check whether or not it increases the discrepancy from the true posterior, compared with the other version. I must admit I always find annoying to see the word bias used in a vague meaning and esp. within a Bayesian setting. All Bayesian methods are biased. End of the story. Quoting our PNAS paper as concluding that ABC model choice is biased is equally misleading: the intended warning represented by the paper was that Bayes factors and posterior probabilities could be quite unrelated with those based on the whole dataset. That the proper choice of summary statistics leads to a consistent model choice shows ABC model choice is not necessarily “biased”… Furthermore, I also fail to understand why the posterior probability of model i should be distributed as a uniform (“If the method is unbiased, the points should fall near the identity line”) when the data is from model i: this is not a p-value but a posterior probability and the posterior probability is not the frequentist coverage…

My overall problem is that, all in all, this is a single if elaborate Monte Carlo study and, as such, it does not carry enough weight to validate an approach that remains highly subjective in the selection of its hyperparameters. Without raising any doubt about an hypothetical “fixing” of those hyperparameters, I think this remains a controlled experiment with simulated data where the true parameters are know and the prior is “true”. This obviously helps in getting better performances.

“With improving numerical methods (…), advances in Monte Carlo techniques and increasing efficiency of likelihood calculations, analyzing rich comparative phylo-geographical models in a full-likelihood Bayesian framework is becoming computationally feasible.” (p.21)

This conclusion of the paper sounds over-optimistic and rather premature. I do not know of any significant advance in computing the observed likelihood for the population genetics models ABC is currently handling. (The SMC algorithm of Bouchard-Côté, Sankaraman and Jordan, 2012, does not apply to Kingman’s coalescent, as far as I can tell.) This is certainly a goal worth pursuing and borrowing strength from multiple techniques cannot hurt, but it remains so far a lofty goal, still beyond our reach… I thus think the major message of the paper is to reinforce our own and earlier calls for caution when interpreting the output of an ABC model choice (p.20), or even of a regular Bayesian analysis, agreeing that we should aim at seeing “a large amount of posterior uncertainty” rather than posterior probability values close to 0 and 1.

robust Bayesian FDR control with Bayes factors [a reply]

Posted in Statistics, University life with tags , , , , on January 17, 2014 by xi'an

(Following my earlier discussion of his paper, Xiaoquan Wen sent me this detailed reply.)

I think it is appropriate to start my response to your comments by introducing a little bit of the background information on my research interest and the project itself: I consider myself as an applied statistician, not a theorist, and I am interested in developing theoretically sound and computationally efficient methods to solve practical problems. The FDR project originated from a practical application in genomics involving hypothesis testing. The details of this particular application can be found in this published paper, and the simulations in the manuscript are also designed for a similar context. In this application, the null model is trivially defined, however there exist finitely many alternative scenarios for each test. We proposed a Bayesian solution that handles this complex setting quite nicely: in brief, we chose to model each possible alternative scenario parametrically, and by taking advantage of Bayesian model averaging, Bayes factor naturally ended up as our test statistic. We had no problem in demonstrating the resulting Bayes factor is much more powerful than the existing approaches, even accounting for the prior (mis-)modeling for Bayes factors. However, in this genomics application, there are potentially tens of thousands of tests need to be simultaneously performed, and FDR control becomes necessary and challenging. Continue reading

Bayesian non-parametrics

Posted in Statistics with tags , , , , , , , , , , , on April 8, 2013 by xi'an

Here is a short discussion I wrote yesterday with Judith Rousseau of a paper by Peter Müller and Riten Mitra to appear in Bayesian Analysis.

“We congratulate the authors for this very pleasant overview of the type of problems that are currently tackled by Bayesian nonparametric inference and for demonstrating how prolific this field has become. We do share the authors viewpoint that many Bayesian nonparametric models allow for more flexible modelling than parametric models and thus capture finer details of the data. BNP can be a good alternative to complex parametric models in the sense that the computations are not necessarily more difficult in Bayesian nonparametric models. However we would like to mitigate the enthusiasm of the authors since, although we believe that Bayesian nonparametric has proved extremely useful and interesting, we think they oversell the “nonparametric side of the Force”! Our main point is that by definition, Bayesian nonparametric is based on prior probabilities that live on infinite dimensional spaces and thus are never completely swamped by the data. It is therefore crucial to understand which (or why!) aspects of the model are strongly influenced by the prior and how.

As an illustration, when looking at Example 1 with the censored zeroth cell, our reaction is that this is a problem with no proper solution, because it is lacking too much information. In other words, unless some parametric structure of the model is known, in which case the zeroth cell is related with the other cells, we see no way to infer about the size of this cell. The outcome produced by the authors is therefore unconvincing to us in that it seems to only reflect upon the prior modelling (α,G*) and not upon the information contained in the data. Now, this prior modelling may be to some extent justified based on side information about the medical phenomenon under study, however its impact on the resulting inference is palatable.

Recently (and even less recently) a few theoretical results have pointed out this very issue. E.g., Diaconis and Freedman (1986) showed that some priors could surprisingly lead to inconsistent posteriors, even though it was later shown that many priors lead to consistent posteriors and often even to optimal asymptotic frequentist estimators, see for instance van der Vaart and van Zanten (2009) and Kruijer et al. (2010). The worry about Bayesian nonparametrics truly appeared when considering (1) asymptotic frequentist properties of semi-parametric procedures; and (2) interpretation of inferential aspects of Bayesian nonparametric procedures. It was shown in various instances that some nonparametric priors which behaved very nicely for the estimation of the whole parameter could have disturbingly suboptimal behaviour for some specific functionals of interest, see for instance Arbel et al. (2013) and Rivoirard and Rousseau (2012). We do not claim here that asymptotics is the answer to everything however bad asymptotic behaviour shows that something wrong is going on and this helps understanding the impact of the prior. These disturbing bad results are an illustration that in these infinite dimensional models the impact of the prior modelling is difficult to evaluate and that although the prior looks very flexible it can in fact be highly informative and/or restrictive for some aspects of the parameter. It would thus be wrong to conclude that every aspect of the parameter is well-recovered because some are. It has been a well-known fact for Bayesian parametric models, leading to extensive research on reference and other types of objective priors. It is even more crucial in the nonparametric world. No (nonparametric) prior can be suited for every inferential aspect and it is important to understand which aspects of the parameter are well-recovered and which ones are not.

We also concur with the authors that Dirichlet mixture priors provide natural clustering mechanisms, but one may question the “natural” label as the resulting clustering is quite unstructured, growing in the number of clusters as the number of observations increases and not incorporating any prior constraint on the “definition” of a cluster, except the one implicit and well-hidden behind the non-parametric prior. In short, it is delicate to assess what is eventually estimated by this clustering methods.

These remarks are not to be taken criticisms of the overall Bayesian nonparametric approach, just the contrary. We simply emphasize (or recall) that there is no such thing as a free lunch and that we need to post the price to pay for potential customers. In these models, this is far from easy and just as far from being completed.”

References

  • Arbel, J., Gayraud, G., and Rousseau, J. (2013). Bayesian adaptive optimal estimation using a sieve prior. Scandinavian Journal of Statistics, to appear.

  • Diaconis, P. and Freedman, D. (1986). On the consistency of Bayes estimates. Ann. Statist., 14:1-26.

  • Kruijer, W., Rousseau, J., and van der Vaart, A. (2010). Adaptive Bayesian density estimation with location-scale mixtures. Electron. J. Stat., 4:1225-1257.

  • Rivoirard, V. and Rousseau, J. (2012). On the Bernstein Von Mises theorem for linear functionals of the density. Ann. Statist., 40:1489-1523.

  • van der Vaart, A. and van Zanten, J. H. (2009). Adaptive Bayesian estimation using a Gaussian random field with inverse Gamma bandwidth. Ann. Statist., 37:2655-2675.

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