Archive for ABC in London

the Frankenstein chronicles

Posted in Statistics with tags , , , , , , , , , , , , , on March 31, 2019 by xi'an

Over a lazy weekend, I watched the TV series The Frankenstein Chronicles, which I found quite remarkable (if definitely Gothic and possibly too gory for some!). Connections with celebrities of (roughly) the time abound: While Mary Shelley makes an appearance in the first season of the series, not only as the writer of the famous novel (already famous in the novel as well) but also as a participant to a deadly experiment that would succeed in the novel (and eventually in the series), Charles Dickens is a constant witness to the unraveling of scary events as Boz the journalist, somewhat running after the facts, William Blake dies in one of the early episodes after painting a series of tarot like cards that eventually explains it all, Ada Lovelace works on the robotic dual of Frankenstein, Robert Peel creates the first police force (which will be called the Bobbies after him!), John Snow’s uncovering of the cholera source as the pump of Broad Street is reinvented with more nefarious reasons, and possibly others. Besides these historical landmarks (!), the story revolves around the corpse trafficking that fed medical schools and plots for many a novel. The (true) Anatomy Act is about to pass to regulate body supply for anatomical purposes and ensues a debate on the end of God that permeates mostly the first season and just a little bit the second season, which is more about State versus Church… The series is not without shortcomings, in particular a rather disconnected plot (which has the appeal of being unpredictable of jumping from one genre to the next) and a repeated proneness of the main character to being a scapegoat, but the reconstitution of London at the time is definitely impressive (although I cannot vouch for its authenticity!). Only the last episode of Season 2 feels a bit short when delivering, by too conveniently tying up all loose threads.

a Bayesian criterion for singular models [discussion]

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

London Docks 12/02/09[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]

approximate Bayesian inference

Posted in Books, pictures, Statistics, Travel, University life with tags , , , , , , , , on March 23, 2016 by xi'an

Maybe it is just a coincidence, but both most recent issues of Bayesian Analysis have an article featuring approximate Bayesian inference. One is by Daniel Add Contact Form Graham and co-authors on Approximate Bayesian Inference for Doubly Robust Estimation, while the other one is by Chris Drovandi and co-authors from QUT on Exact and Approximate Bayesian Inference for Low Integer-Valued Time Series Models with Intractable Likelihoods. The first paper has little connection with ABC. Even though it (a) uses a lot of three letter acronyms [which does not help with speed reading] and (b) relies on moment based and propensity score models. Instead, it relies on Bayesian bootstrap, which suddenly seems to me to be rather connected with empirical likelihood! Except the weights are estimated via a Dirichlet prior instead of being optimised. The approximation lies in using the bootstrap to derive a posterior predictive. I did not spot any assessment or control of the approximation effect in the paper.

“Note that we are always using the full data so avoiding the need to choose a summary statistic” (p.326)

The second paper connects pMCMC with ABC. Plus pseudo-marginals on the side! And even simplified reversible jump MCMC!!! I am far from certain I got every point of the paper, though, especially the notion of dimension reduction associated with this version of reversible jump MCMC. It may mean that latent variables are integrated out in approximate (marginalised) likelihoods [as explicated in Andrieu and Roberts (2009)].

“The difference with the common ABC approach is that we match on observations one-at-a-time” (p.328)

The model that the authors study is an integer value time-series, like the INAR(p) model. Which integer support allows for a non-zero probability of exact matching between simulated and observed data. One-at-a-time as indicated in the above quote. And integer valued tolerances like ε=1 otherwise. In the case auxiliary variables are necessary, the authors resort to the alive particle filter of Jasra et al. (2013), which main point is to produce an unbiased estimate of the (possibly approximate) likelihood, to be exploited by pseudo-marginal techniques. However, unbiasedness sounds less compelling when moving to approximate methods, as illustrated by the subsequent suggestion to use a more stable estimate of the log-likelihood. In fact, when the tolerance ε is positive, the pMCMC acceptance probability looks quite close to an ABC-MCMC probability when relying on several pseudo-data simulations. Which is unbiased for the “right” approximate target. A fact that may actually holds for all ABC algorithms. One quite interesting aspect of the paper is its reflection about the advantage of pseudo-marginal techniques for RJMCMC algorithms since they allow for trans-dimension moves to be simplified, as they consider marginals on the space of interest. Up to this day, I had not realised Andrieu and Roberts (2009) had a section on this aspect… I am still unclear about the derivation of the posterior probabilities of the models under comparison, unless it is a byproduct of the RJMCMC algorithm. A last point is that, for some of the Markov models used in the paper, the pseudo observations can be produced as a random one-time move away from the current true observation, which makes life much easier for ABC and explain why exact simulations can sometimes be produced. (A side note: the authors mention on p.326 that EP is only applicable when the posterior is from an exponential family, while my understanding is that it uses an exponential family to approximate the true posterior.)

ABC in Sydney, July 3-4, 2014!!!

Posted in pictures, Statistics, Travel, University life, Wines with tags , , , , , , , , , , , , , , on February 12, 2014 by xi'an

Sydney Opera from Sydney Harbour Bridge, Sydney, July 14, 2012After ABC in Paris in 2009, ABC in London in 2011, and ABC in Roma last year, things are accelerating since there will be—as I just learned—  an ABC in Sydney next July (not June as I originally typed, thanks Robin!). The workshop on the current developments of ABC methodology thus leaves Europe to go down-under and to take advantage of the IMS Meeting in Sydney on July 7-10, 2014. Hopefully, “ABC in…” will continue its tour of European capitals in 2015! To keep up with an unbroken sequence of free workshops, Scott Sisson has managed to find support so that attendance is free of charge (free as in “no registration fee at all”!) but you do need to register as space is limited. While I would love to visit UNSW and Sydney once again and attend the workshop, I will not, getting ready for Cancún and our ABC short course there.

fie on fee frenzy!

Posted in Mountains, Statistics, Travel, University life with tags , , , , , , , , , , , , on June 11, 2013 by xi'an

London by Delta, Dec. 14, 2011In the past years, I noticed a clear inflation on conference fees, inflation that I feel unjustified… I already mentioned the huge $720 fees for the Winter Simulation Conference (WSC 2012), which were certainly not all due to the heating bill! Even conferences held by and in universities or societies seem to face the same doom: to stick to conferences I will attend—and do support, to the point of being directly or indirectly involved—, take for instance Bayes 250 in London (RSS Headquarters), £135, Bayes 250 at Duke, $190, both one day-long, and O-Bayes 2013, also at Duke, $480 (in par with JSM fees)… While those later conferences include side “benefits” like meals and banquet, the amount remains large absolutive. Too large. And prohibitive for participants from less-favoured countries (possibly including speakers themselves in the case of O-Bayes 2013). And also counter-productive in the case of both Bayes 250 conferences since we want to get together to celebrate two and a half centuries of Bayesian statistics. Since most of the talks there will be partly commemorative, rather than on the brink of research, I fear some people may have to make a choice to allocate their meagre research funds to other conferences. And I do not understand why universities now consider organising meetings as a source of income rather than as a natural part of their goals.

Now, you may ask, and what about MCMski on which I have more than a modicum of control..?! Well, the sole cost there is renting the conference centre in Chamonix, which is the only place I knew where a large conference could be held. Apart from that, no frill! The coffee breaks will be few and frugal, there will be no free lunch or breakfast or banquet, and no one will get a free entry or a paid invitation. As a result, the registration fee is only 170€ for three days (plus a free satellite meeting the next day), an amount computed on an expected number of participants of 150 and which could lead me to pay the deficit from my own research grants in case I am wrong.  (And may I recall the “ABC in…” series, which has been free of fees so far!)

My point, overall, is that we should aim at more frugal meetings, in order to attract larger and more diverse crowds (even though fees are only part of the equation, lodging and travelling can be managed to some extent as long as the workshop is not in too an exotic location).

efficient learning in ABC

Posted in Statistics with tags , , , , , , on October 11, 2012 by xi'an

Jean-Michel Marin just posted on arXiv a joint paper of ours, Efficient learning in ABC algorithms. This paper, to which elaboration [if not redaction] I contributed to, is one of the chapters of Mohammed Sedki’s thesis. (Mohammed is about to defend this thesis, currently with reviewers. A preliminary version of this paper was presented at ABC in London and it is in revision with Statistics and Computing. Hence missing the special issue!)

The paper builds on the sequential ABC scheme of Del Moral et al. (2012), already discussed in this post, where the tolerance level at each step is adapted from the previous iterations as a quantile of the distances. (The quantile level is based on a current effective sample size.) In a “systematic” step, the particles that are closest to the observations are preserved and duplicated, while those farther away are sampled randomly. The resulting population of particles is then perturbed by an adaptive (random walk) kernel and the algorithm stops when the tolerance level does not decrease any longer or when the acceptance rate of the Metropolis step is too low. Pierre Pudlo and Mohammed Sedki experimented a parallel implementation of the algorithm, with an almost linear improvement in the number of cores. It is less clear the same would work on a GPU because of the communication requirements. Overall, the new algorithm brings a significant improvement in computing time when compared with earlier versions, mainly because the number of simulations from the model is minimised. (It was tested on a rather complex population scenario retracing the invasion of honeybees in Europe (in connection with the previous post!)

ABC in Roma, May 30-31, 2013!!!

Posted in pictures, Statistics, Travel, University life, Wines with tags , , , , , , on July 24, 2012 by xi'an

After ABC in Paris in 2009 and ABC in London in 2013, the ABC workshop on the current developments of ABC methodology continues its tour of European capitals! It will take place next year in Rome over two days to allow for travel from Paris and London (no EuRomaStar yet!). ABC in Rome is organised by Brunero Liseo and his colleagues at Roma 1 and Roma 3 Università, and sponsored by La Sapienza Università di Roma. Following the previous meetings, in Paris and London, the field still sees rapid methodology progress and an increased number of applications in a wider range of scientific areas, as no doubt readers of the ‘Og are aware! ABC in Rome (ABCiR) will hopefully bring together leading researchers in the field, with focus on

  •  applications of ABC to real world problems
  • recent computational advances in ABC
  • comparative efficiency of ABC methods with respect to alternative methodologies
  • model selection and model checking in the ABC framework.

As in the previous meetings, attendance is free of charge (free as in “no registration fee at all”!) but you do need to register as space is strictly limited. (For ABC in London, the waiting list was more than 100 persons long…) If you wish to present a poster then please email the organisation committee with a brief abstract. We particularly encourage posters from young participants and posters detailing recent software implementations of ABC methods and computational advance.