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Innovative Methodology

Improved Spike-Sorting By Modeling Firing Statistics and Burst-Dependent Spike Amplitude Attenuation: A Markov Chain Monte Carlo Approach

¹Laboratoire de Physiologie Cérébrale, Centre National de la Recherche Scientifique (CNRS) Unité Mixte de Recherche (UMR) 8118, Université René Descartes, 75006, Paris; ²Laboratoire de Physique Théorique des Liquides, CNRS UMR 7600, Université Pierre et Marie Curie, 75252 Paris Cedex 05; and ³Laboratoire d'Analyse et de Mathématiques Appliquées, CNRS UMR 8050, Université de Marne la Vallée, Batiment Copernic, Cité Descartes, 77454 Marne la Vallée Cedex, France

Submitted 10 March 2003; accepted in final form 14 January 2004

	ABSTRACT

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
Spike-sorting techniques attempt to classify a series of noisy electrical waveforms according to the identity of the neurons that generated them. Existing techniques perform this classification ignoring several properties of actual neurons that can ultimately improve classification performance. In this study, we propose a more realistic spike train generation model. It incorporates both a description of "nontrivial" (i.e., non-Poisson) neuronal discharge statistics and a description of spike waveform dynamics (e.g., the events amplitude decays for short interspike intervals). We show that this spike train generation model is analogous to a one-dimensional Potts spin-glass model. We can therefore tailor to our particular case the computational methods that have been developed in fields where Potts models are extensively used, including statistical physics and image restoration. These methods are based on the construction of a Markov chain in the space of model parameters and spike train configurations, where a configuration is defined by specifying a neuron of origin for each spike. This Markov chain is built such that its unique stationary density is the posterior density of model parameters and configurations given the observed data. A Monte Carlo simulation of the Markov chain is then used to estimate the posterior density. We illustrate the way to build the transition matrix of the Markov chain with a simple, but realistic, model for data generation. We use simulated data to illustrate the performance of the method and to show that this approach can easily cope with neurons firing doublets of spikes and/or generating spikes with highly dynamic waveforms. The method cannot automatically find the "correct" number of neurons in the data. User input is required for this important problem and we illustrate how this can be done. We finally discuss further developments of the method.

	INTRODUCTION

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
The study of neuronal populations activity is one of the main experimental challenges of contemporary neuroscience. Among the techniques that have been developed and used to monitor populations of neurons, the multielectrodes extracellular recordings, albeit one of the oldest, still stands as one of the methods of choice. It is relatively simple and inexpensive to implement (especially when compared to imaging techniques) and it can potentially give access to the activity of many neurons with a fine time resolution (below the ms). The full exploitation of this method nevertheless requires some basic problems to be solved, among which stands prominently the spike-sorting problem. The raw data collected by extracellular recordings are of course corrupted by some recording noise but, more important, are a mixture of activities from different neurons. This means that even in good situations where action potentials or spikes can be unambiguously distinguished from background noise, the experimentalist is left with the intricate problem of finding out how many neurons are in fact contributing to the recorded data and, for each spike, which is the neuron of origin. This in essence is the spike-sorting problem. Extracellular recordings being an old method, the problem has been recognized as such for a long time and there is already a fairly long history of proposed solutions (Lewicki 1998). Nevertheless it seems to us that none of the available methods makes full use of the information present in the data. They consider mainly or exclusively the information provided by the waveform of the individual spikes and they neglect that provided by their occurrence time. Yet, the importance of the spikes occurrence times shows up in two ways.

1) First, the sequence of spike occurrence times emitted by a neuron has well-known features like the presence of a refractory period (e.g., after a spike has been emitted no other spike will be emitted by the same neuron for a period whose duration varies from 2 to 10 ms). Other features include an interspike interval (ISI) probability density, which is often unimodal and skewed (it rises "fast" and decays "slowly"). Although some methods make use of the presence of a refractory period (Fee et al. 1996a; Harris et al. 2000), none makes use of the full ISI density. This amounts to throwing useful information away, for if event 101 has been attributed to neuron 3 we not only know that no other spike can arise from neuron 3 for, say, the next 5 ms, but we also know that the next spike of this neuron is likely to occur with an ISI between 10 and 20 ms. As pointed out by Lewicki (1994), any spike-sorting method that would include an estimate of the ISI density of the different neurons should produce better classification performances.

2) Second, the spike waveform generated by a given neuron often depends on the time elapsed since its last spike (Quirk et al. 1999). This nonstationarity of spike waveform will worsen the classification reliability of methods that assume stationary waveform (Lewicki 1994; Pouzat et al. 2002). Other methods (e.g., Gray et al. 1995 for a "manual" method and Fee et al. 1996a for an automatic one) assume a continuity of the cluster of events originating from a given neuron when represented in a "feature space" [the peak amplitude, valley(s) amplitude(s), half-width are examples of such features] and make use of this property to classify nonstationary spikes. Although it is clear that the latter methods will in general outperform the former, we think that some cases could arise where none of them would work. In particular, doublet-firing cells could generate well-separated clusters in feature space and these clusters would not be put together by presently available methods.

These considerations motivated us to look for a new method that would have a built-in capability to take into account both the ISI density of the neurons and their spike waveform dynamics. However, there are two additional issues with spike-sorting that, we think, are worth considering.

1) "Hard" versus "soft" spike-sorting. Most spike-sorting methods, including the ones where users actually perform the clustering and some automatic methods like the one of Fee et al. (1996a), generate "hard" classification. That is, each recorded event is attributed "fully" to one of the K neurons contributing to the data. Methods based on probabilistic models (Lewicki 1994; Nguyen et al. 2003; Pouzat et al. 2002; Sahani 1999) deal with this issue differently: each event is given a probability to have been generated by each of the K neurons (e.g., one gets results like: event 100 was generated by neuron 1 with probability 0.75 and by neuron 2 with probability 0.25). We will refer to this kind of classification as "soft" classification. Users of the latter methods are often not aware of their soft classification aspect because what one does for subsequent processing like estimating the ISI density of a given neuron or computing the cross-correlogram of spike trains from 2 different neurons, is to force the soft classification into the most likely one (in the example above we would force the classification of event 100 into neuron 1). By doing so, however, the analyst introduces a bias in the estimates. It seems therefore interesting to find a way to keep the soft classification aspect of the model-based methods when computing estimates.

2) Confidence intervals on model parameters. Methods based on an explicit model for data generation could in principle generate confidence intervals for their model parameters, although this was never done. However, the values of model parameters do influence the classification and a source of bias could be reduced by including information about uncertainty of model parameters values.

Based on these considerations we have developed a semiautomatic method that takes spike waveform dynamics and ISI densities into account, which produces soft classification and allows the user to make full use of it, and which generates a full posterior density for the model parameters and confidence intervals. Our method uses a probabilistic model for data generation where the label (i.e., neuron of origin) of a given spike at a given time depends on the label of other spikes occurring just before or just after. We will call configuration the specification of a label for each spike in the train. It will become clear that with this model the spike-sorting problem is analogous to an image restoration problem where a picture (i.e., a set of pixel values) that has been generated by a real object and corrupted by some noise is given. The problem is to find out what the actual pixel value was, given the noise properties and the known correlation properties of noise free images. The case of spike sorting is equivalent to a one-dimensional "pixel sequence" where the pixel value is replaced by the label of the spike. More generally it is a special case of a Potts spin-glass model encountered in statistical physics (Newman and Barkema 1999; Wu 1982). Thanks to this analogy solutions developed to study Potts models (Landau and Binder 2000; Newman and Barkema 1999) and to solve the image restoration problem (Geman and Geman 1984) can be tailored to the spike-sorting problem. These solutions rely on a Markov chain that has the posterior density of model parameters and configurations given the observed data as its unique stationary density. This Markov Chain is stochastically simulated on the computer. This general class of methods has been used for 50 years by physicists (Metropolis et al. 1953) who call it Dynamic Monte Carlo and for 20 years by statisticians (Fishman 1996; Geman and Geman 1984; Liu 2001; Robert and Cassela 1999) who call it Markov Chain Monte Carlo (MCMC).

Our purpose in this paper is 2-fold: First to demonstrate that a particular MCMC method allows incorporation of more realistic data generation models and by doing so to perform reliable spike-sorting even in difficult cases (e.g., in the presence of neurons generating 2 well-separated clusters in feature space). Second, to explain how and why this method works, thereby allowing its users to judge the quality of the produced classification, to improve the algorithm implementing it, and to adapt it to their specific situations. Our method does not yet provide an automatic estimate of the number of neurons present in the data, although we illustrate how this critical problem can be addressed by the user.

	METHODS

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
Data generation model

In the present manuscript we will make the following assumptions about data generation.

1. The firing statistics of each neuron are fully described by a time-independent interspike interval density. That is, the sequence of spike times from a given neuron is a realization of a homogeneous renewal point process (Johnson 1996).
   
2. The spike amplitudes generated by each neuron depend on the elapsed time since the previous spike of this neuron.
   
3. The measured spike amplitudes are corrupted by a Gaussian white noise, which sums linearly with the spikes and is statistically independent of them.
   

These assumptions have been chosen to show as simply as possible the implementation of our method and should not be taken as intrinsic limitations of this method. They constitute, moreover, a fairly good first approximation of real data in our experience. More sophisticated models where the next ISI of a neuron depends on the value of the former one (Johnson et al. 1986) could easily be included. In the same vein, models where the amplitude of a spike depends on several of the former ISIs could be considered. The Gaussian white noise assumption means that the analyst has "whitened" the noise before starting the spike-sorting procedure (Pouzat et al. 2002).

INTERSPIKE INTERVAL DENSITY. We will use in this paper a log-Normal density for the ISIs. This density looks like a good next guess, after the exponential density, when one tries to fit empirical ISI densities. It is unimodal, exhibits a refractory period, rises "fast," and decays "slowly." The version of the log-Normal density we are using depends on 2 parameters: a dimensionless shape parameter and a scale parameter s measured in seconds. The probability density for a realization of a log-Normal random variable I with parameters and s to have the value i is

(1)

In what follows we will use a shorter notation

meaning that i is a realization of a log-Normal random variable with parameters and s. A discussion of the properties of the log-Normal distribution can be found in the e-Handbook of Statistical Methods.

SPIKE AMPLITUDE DYNAMICS. We will consider events described by their occurrence time and their peak amplitude measured on one or several recording sites. This makes the equations more compact, but full waveforms can be accommodated if necessary. Following Fee et al. (1996b) we will describe the dependence of the amplitude on the ISI by an exponential relaxation

(2)

where i is the ISI, is the inverse of the relaxation time constant (measured in 1/s), P is the vector of the maximal amplitude of the event on each recording site (i.e., this is the amplitude observed when i >> ^-1), and is the maximal modulation [i.e., for i << ^-1 we have A(i) P(1 - )]. It is clear from this equation that the amplitudes of events generated by a given neuron are modulated by the same relative amount on each recording site. This is one of the key features of modulation observed experimentally (Gray et al. 1995).

To keep equations as compact as possible we will assume that the amplitudes of the recorded events on each recording site have been normalized by the SD of the noise on the corresponding site. A and P in Eq. 2 are therefore dimensionless and the noise SD equals 1. Combining Eq. 2 with our third model hypothesis (independent Gaussian noise corruption) we get for the density of the amplitude vector a of a given neuron conditioned on the ISI value i

(3)

where n_s is the number of recording sites and ||v|| stands for the Euclidean norm of v. We will sometimes express Eq. 3 in a more compact form

(4)

where Norm (m, v) stands for a normal distribution with mean m and whose covariance matrix is diagonal with diagonal values equal to v.

We get the density of the couple (i, a) by combining Eq. 1 with Eq. 3

(5)

NOTATIONS FOR THE COMPLETE MODEL AND THE DATA. We now have, for each neuron in the model, 2 parameters ( and s) to specify the ISI density and 2 + number of recording sites parameters (e.g., for tetrode recordings: P₁, P₂, P₃, P₄, , ) to specify the amplitude density conditioned on the ISI value (Eq. 3). That is, in the case of tetrode recording: 8 parameters per neuron. In the sequel we will use the symbol to refer to the complete set of parameters specifying our model. That is for a model with K neurons applied to tetrode recordings

(6)

We will assume that our data sample is of size N and that each element j of the sample is fully specified by its occurrence time t_j and its peak amplitude(s), a_j,1, a_j,2, a_j,3, a_j,4 (for tetrode recording). We will use Y to refer to the full data sample, that is

(7)

Data augmentation, configurations, and likelihood function

DATA AUGMENTATION AND CONFIGURATIONS. Until now we have specified our model and our data representation but what we are really interested in is to find the neuron of origin of each individual spike, or more precisely, for a model with K neurons, the probability for each spike to have been generated by each of the K neurons (remember that we are doing soft classification). To do that we will associate with each spike j a latent variable, c_j {1, 2,..., K}. When c_j has value 2, that means that spike j has been "attributed" to neuron 2 of the model. We will use C to refer to the set of N variables c

(8)

For a data sample of size N and a model with K neurons we have K^N different C values. We will call configuration a specific C value. That is, a configuration is a N-dimensional vector whose components take value in {1, 2,..., K} [e.g., (1,..., 1), (K,..., K), are 2 configurations]. In the statistical literature, this procedure of "making the data more informative" by adding a latent variable is called data augmentation (Liu 2001; Robert and Casella 1999).

LIKELIHOOD FUNCTION. Because C has been introduced it is pretty straightforward to compute the likelihood function of the "augmented" data, given specific values of : L(Y, C|). We remind the reader here that the likelihood function is proportional to the probability density of the (augmented) data given the parameters. Figure 1 illustrates how this is done on a simple case with 2 units in the model (K = 2) and a single recording site.

View larger version (11K): [in this window] [in a new window]   FIG. 1. Likelihood computation. A: snapshot of spikes 11 to 15 of a train. cj values are shown at the top. Spike amplitudes are given (aj) as well as the occurrence times (tj). B: spikes from units 1 are shown alone. C: spikes from units 2 are shown alone.

(9)

where Y_l is the subsample of Y for which the components c_j of C are equal to l. The L(Y_l, C|) themselves are product of terms like Eq. 5 In the case illustrated on Fig. 1 we will have for the first train

(10)

where t_1,former stands for the time of the former spike attributed to neuron 1 in configuration C and t_1,next stands for the next spike attributed to neuron 1 in configuration C. If there are N₁ spikes attributed to neuron 1 then, using periodic boundary conditions (see Parameter-specific transition kernels), there will be N₁ terms in L(Y₁, C|). For the second neuron of Fig. 1 we obtain

(11)

The reader can see that for any configuration Eq. 9 involves the computation of N terms.

Posterior and prior densities, a combinatorial explosion

POSTERIOR DENSITY. Now that we know how to compute the likelihood function we can write the expression of the posterior density of model parameters and configuration given the data using Bayes formula

(12)

where _prior() is the prior density of the parameters and

(13)

where Z is the probability of the data. It is clear that if we manage to compute or estimate _post(, C|Y) we will have done our job, for then the soft classification is given by the marginal density of C

(14)

The reader will note that this posterior marginal density includes all the information available about , that is, the uncertainty on the parameters is automatically taken into account.

PRIOR DENSITY. We will assume here that several thousand spikes have been collected and that we perform the analysis by taking chunks of data, say the first 5,000 spikes, then spikes 5,001 to 1,000 and so on. For the first chunk we will assume we know "little" a priori and that the joint prior density _prior() can be written as a product of the densities for each component of , that is

where we are assuming that 4 recording sites have been used. We will further assume that our signal to noise ratio is not better 20 (a rather optimistic value), that our spikes are positive, and therefore the (P_{q,1· · · 4}) are null below 0 and above +20 (remember we are working with normalized amplitudes). We will reflect our absence of prior knowledge about the amplitudes by taking a uniform distribution between 0 and +20. The value reported by Fee et al. (1996b) is 45.5 s^-1. must, moreover, be smaller than , so we took a prior density uniform between 10 and 200 s^-1. must be 1 (the amplitude of a spike from a given neuron on a given recording site does not change sign) and 0 (spikes do not become larger upon short ISI), so we used a uniform density between 0.1 and 0.9 for . An inspection of the effect of the shape parameter on the ISI density is enough to convince an experienced neurophysiologist that empirical unimodal ISI densities from well-isolated neurons will have [0.1, 2]. We therefore took a prior density uniform between 0.1 and 2 for . The same considerations led us to take a uniform prior density between 0.005 and 0.5 for s.

When one analyzes the following data chunks, say spikes 5,001 to 1,000, the best strategy is to take as a prior density for the posterior density from the previous chunk

One has therefore a direct way to track parameters drifts during an experiment.

A COMBINATORIAL EXPLOSION. We have so far avoided considering the normalizing constant Z of Eq. 12, whose expression is given by Eq. 13. A close examination of this equation shows that it involves a multidimensional integration in a rather highly dimensional space (e.g., for a model with 10 neurons and data from tetrode recordings, we have 80 components in ) and a summation over every possible configuration, that is a sum of K^N terms. That is really a lot, to say the least, in any realistic case (say, N = 1,000 and K = 10)! One can wonder how we will manage to compute Z and therefore implement our approach. Fortunately, we do not really need to compute it, as will now be explained.

The Markov Chain Monte Carlo method

AN ANALOGY WITH THE POTTS MODEL IN STATISTICAL PHYSICS. Let us define

(15)

that is, e(, C|Y) is an unnormalized version of _post(, C|Y), the posterior density, and

(16)

Then Eq. 12 can be rewritten as

(17)

with = (1/kT) = 1. If one interprets E as an energy, as an "inverse temperature," Z as a partition function, one sees that the posterior density, _post(, C|Y), is the canonical density (or canonical distribution) used in statistical physics (Landau and Binder 2000; Newman and Barkema 1999). A closer examination of Eqs. 10 and 11 shows that, from a statistical physics viewpoint, the (log of the) likelihood function describes "interactions" between spikes attributed to the same neuron. If one goes a step further and replaces "spikes" by "atoms on crystal nodes" and "neuron of origin" by "spin orientation," ones sees that our problem is analogous to what physicists call a one-dimensional Potts model. The typical Potts model is a system of interacting spins where each spin interacts only with its nearest neighbors and only if they have the same spin value (Wu 1982). In our case, spins (that is, spikes) interact with the former and next spins with the same value regardless of the distance (time interval) at which they are located. Moreover the "interaction" strength is a random variable that makes our system analogous to a spin glass (Binder and Young 1986; Landau and Binder 2000; Newman and Barkema 1999) and the amplitude contribution (Eq. 3) gives rise to an energy term analogous to an interaction between a spin and a random magnetic field (Newman and Barkema 1999). To be complete we should say that physicists consider that is given, then Eq. 17 gives the probability to find the "crystal" in a specific configuration C.

The point of this analogy will now become clear. Physicists are interested in estimating expected values (which are what they can experimentally measure) of functions that can be easily computed for each configuration. For instance, they want to compute the expected energy of the "crystal"

(18)

To do that they obviously have to deal with the combinatorial explosion we alluded to. The idea of their solution is to draw configurations: C⁽¹⁾, C⁽²⁾,..., C^(m), from the "target" distribution: _post(, C|Y). Then an estimate of is the empirical average

(19)

In other words, they use a Monte Carlo (MC) method (Landau and Binder 2000; Newman and Barkema 1999). The problem becomes therefore to generate the draws [C⁽¹⁾, C⁽²⁾,..., C^(m)].

A SOLUTION BASED ON A MARKOV CHAIN. For reasons discussed in detail by Newman and Barkema (1999) and Liu (2001) the draws cannot be generated in practice by "direct" methods like rejection sampling (Fishman 1996; Liu 2001; Robert and Casella 1999). By "direct" we mean here that the C^(t) are independent and identically distributed draws from _post(, C|Y) for a given value. Instead we must have recourse to an artificial dynamics generating a sequence of correlated draws. This sequence of draws is in practice the realization of a Markov chain. That means that in the general case, where we are interested in getting both and C, we will use a transition kernel: T[^(t), C^(t)|^(t-1), C^(t-1)], and simulate the chain starting from [⁽⁰⁾, C⁽⁰⁾], chosen such that _post[⁽⁰⁾, C⁽⁰⁾|Y] > 0. Moreover, we will build T such that the chain converges to a unique stationary distribution given by _post(, C|Y). The estimator Eq. 19 of the expected value Eq. 18 is then still correct even though the successive [^(t), C^(t)] are correlated; we just need to be cautious when we compute the variance of the estimator (Janke 2002; Sokal 1989; Empirical averages and SDs). By still correct we formally mean that the following equality holds

(20)

We give in The Metropolis–Hastings algorithm of the APPENDIX a general presentation and justification of the procedure used to build the transition kernel T: the Metropolis–Hastings (MH) algorithm. We then continue in Parameter-specific transition kernels with an account of the specific kernel we used for the analysis presented in this paper.

Empirical averages and SDs

As explained at the beginning of Slow relaxation and REM, once our Markov chain has reached equilibrium, or at least once its behavior is compatible with the equilibrium regime, we can estimate values of parameters of interest as well as errors on these estimates. The estimator of the probability for a given spike, say the 100th, to originate from a given neuron, say the second, is straightforward to obtain. If we assume that we discard the first N_D = 15,000 on a total of N_T = 20,000 iterations we have

(21)

where I_q is the indicator function defined by

(22)

In a similar way, the estimate of the expected value of the maximal peak amplitude of the first neuron on the first recording site is

To obtain the error on such estimators we have to keep in mind that the successive states [^(t), C^(t)] generated by the algorithm are correlated. Thus, we cannot use the empirical variance divided by the number of generated states to estimate the error (Janke 2002; Sokal 1989). As explained in detail by Janke (2002) we have to compute for each parameter _i of the model the normalized autocorrelation function (ACF), _norm(l; _i), defined by

(23)

Then we compute the integrated autocorrelation time, _autoco(_i)

(24)

where L is the lag at which starts oscillating around 0. Using an empirical variance, ²(_i) of parameter _i, defined in the usual way

(25)

Our estimate of the variance, Var of becomes

(26)

The first consequence of the autocorrelation of the states of the chain is therefore to reduce the effective sample size by a factor of 2_autoco(_i). This gives us a first quantitative element on which different algorithms can be compared (as explained in The Metropolis–Hastings algorithm of the APPENDIX, the MH algorithm does in fact give us a lot of freedom on the choice of proposal transition kernels). It is clear that the faster the autocorrelation functions of the parameters fall to zero, the greater the statistical efficiency of the algorithm. The other quantitative element we want to consider is the computational time, _cpu, required to perform one MC step of the algorithm. One could for instance imagine that a new sophisticated proposal transition kernel allows us to reduce the largest _autoco of our standard algorithm by a factor of 10, but at the expense of an increase of _cpu by a factor of 100. Globally the new algorithm would be 10 times less efficient than the original one. What we want to keep as small as possible is therefore the product _autococpu. With this efficiency criterion in mind, the replica exchange method described in Slow relaxation and REM becomes even more attractive.

Initial guess

MCMC methods are iterative methods that must be started from an initial guess. We chose randomly with a uniform probability 1/N as many actual events as neurons in the model (K). That gave us our initial guesses for the P_q,i. was set to _min for each neuron. All the other parameters were randomly drawn from their prior distribution. The initial configuration was generated by labeling each individual spike with one of the K possible labels with a probability 1/K for each label (this is the = 0 initial condition used in statistical physics).

Data simulation

We have tried to make the simulated tetrode data set used in this paper a challenging one. First the noise corruption was generated from a Student's t density with 4 degrees of freedom instead of a Gaussian density as assumed by our model (Eq. 3). Second, among the 6 simulated neurons only 3 had an actual log-Normal ISI density. Among the 3 others, one had a gamma density

(27)

where s is a scale parameter and is a shape parameter. One neuron was a doublet-generating neuron and the third one had a truncated log-Normal ISI density. The truncation of the latter came from the fact that 18% of the events it generated were below the "detection threshold," which would have been at 3 noise SD. The doublet-generating neuron was obtained with 2 log-Normal densities, one giving rise to short ISIs (10 ms) the other one to "long" ones (60 ms); the neuron was moreover switching from the short, respectively the long, ISI generating state to the long, respectively the short, ISI generating state after each spike, giving rise to a strongly bimodal ISI density (Fig. 9C). The mean ISI of these 6 neurons ranged from 10 to 210 ms (Table 1). A "noise" neuron was moreover added to mimic the collective effect of many neurons located far away from the recording sites, which would therefore not be detected most of the time. This "noise" neuron was obtained by generating first a sequence of time points exponentially distributed. A random amplitude given by the background noise density (t density with 4 degrees of freedom; see above) was associated to each point and the point was kept in the "data" sample only if its amplitude exceeded the detection threshold (3 noise SD) on at least one of the 4 recording sites. The initial exponential density for this "noise" neuron was set such that a preset fraction of events in the final data sample would arise from it (in that case 15%, or 758 events on a total of 5,058 events). Details on the pseudorandom number generators required to simulate the data can be found in Random number generators for the simulated data of the APPENDIX.

View larger version (40K): [in this window] [in a new window]   FIG. 9. Comparison between the exact ISI histograms and their estimates. Exact histograms are displayed in gray, their estimates from the first period with crosses and from the second period with circles. SD for the estimated values is always smaller than the size of the symbols.

Codes were written in C and are freely available under the Gnu Public License at our web site: http://www.biomedicale.univ-paris5.fr/physcerv/Spike-0-Matic.html. The free software Scilab (http://www-rocq.inria.fr/scilab/) was used to generate output plots as well as the graphical user interface, which comes with our release of the routines. The GNU Scientific Library (GSL: http://sources.redhat.com/gs1/) was used for vector and matrix manipulation routines and (pseudo)random number generators (Uniform, Normal, log-Normal, Gamma). The GSL implementation of the MT19937 generator of Matsumoto and Nishimura (1998) was used. This generator has a period of 2^19,937 - 1. Because the code requires a lot of exponential and logarithm evaluations, the exponential and logarithm functions were tabulated and stored in memory, which reduced the computation time by 30%. Codes were compiled and run on a PC laptop computer (Pentium IV with CPU speed of 1.6 GHz, 256 MB RAM) running Linux. The gcc compiler (http://www.gnu.org/software/gcc/gcc.htm) version 3.2 was used.

	RESULTS

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
Data properties

The performances of our algorithm will be illustrated in this paper with a simulated data set. The data are supposed to come from tetrode recordings. Six neurons are present in the data together with a "noise" neuron supposed to mimic events rarely detected from neurons far away from the tetrode's recording sites. The data set has been made challenging for the algorithm by producing systematic deviations with respect to our model assumptions (Data simulation). This data set is supposed to come from 15 s of recording during which 5,058 events were detected. Wilson plots as they would appear to the user are shown on Fig. 2A. Figure 2B shows the same data with colors corresponding to the neuron of origin. The parameters used to simulate each individual neuron are given in Table 1. The reader can easily see that, whereas one cluster has very few points (purple cluster on Fig. 2B), others have a lot (brown, green, and blue clusters). Indeed, the smallest cluster contains 73 events, whereas the biggest contains 1,474 events. Some clusters are very elongated (e.g., green and red clusters) and one neuron, the doublet-generating neuron, even gives rise to 2 well-separated clusters (red). The "noise" neuron cluster (black) is much more spread out than the others. Two cluster pairs always exhibit an overlap (the yellow-black and the green-red pairs).

View larger version (26K): [in this window] [in a new window]   FIG. 2. Wilson plots showing about 25% of the sample (1,215 events). On each plot the scale bars meet at amplitude (0, 0) and are of size 2 (in units of noise SD). A: data as the analyst would see them before starting the spike-sorting procedure. B: same data sorted with the known neuron of origin encoded by the color: neuron 1 (blue), neuron 2 (green), neuron 3 (red), neuron 4 (brown), neuron 5 (purple), neuron 6 (yellow), noise events (black).

View larger version (22K): [in this window] [in a new window]   FIG. 3. Noise properties. A1: amplitudes of the events generated by neuron 2 (green on Fig. 2B) on the second recording site are plotted against the corresponding interspike interval (ISI). Ideal amplitude relaxation curve (Eq. 2, with P2 = 15, = 0.8, = 100) is shown in gray. Abscissa scale in s. A2, residuals: actual amplitude - ideal amplitude. B1: histogram of the residual values (broken line), ideal density from which these residuals have been generated (gray, t density with an SD = 1 and 4 degrees of freedom) and the density assumed by our model (black, Gaussian density with an SD = 1). B2: t (gray) and Gaussian (black) densities on a log scale showing the heavier tails of the t density.

We want to illustrate here some basic features of our algorithm dynamics as well as how model comparison can be done. In its present form, our approach requires the user to decide what is the proper model (i.e., the proper number of neurons). This decision is based on the examination of "empirical" features associated with the different models. To obtain these empirical features, we have to perform the same computation on the different models, although the computation time increases with the number of events considered. It is therefore a good idea to start with a reduced sample that contains enough events for the model comparison to be done and little enough for the computation to run quickly. In this section we will work with the first 3 s of data (1,017 events), which represent approximately 20% of the total sample (5,058 events). We will focus here on a model with 7 neurons, but a similar study was performed with models containing from 4 to 10 neurons.

EVIDENCE FOR META-STABLE STATES. To explore a model with a given number of neurons we start by simulating 10 different and independent realizations of a Markov chain. The way random initial guesses for these different realizations are obtained is explained in Initial guess. For each realization, one Monte Carlo (MC) step consists in an update of the label of each of the 1,017 spikes (SPIKE LABEL TRANSITION MATRIX), an update of the amplitude parameters (P, , ) (AMPLITUDE PARAMETERS TRANSITION KERNELS) and of the 2 parameters of the ISI density, s and (SCALE PARAMETER TRANSITION KERNEL & SHAPE PARAMETER TRANSITION KERNEL) for each neuron. As explained in AMPLITUDE PARAMETERS TRANSITION KERNELS, the amplitude parameters are generated with a piecewise linear approximation of the corresponding posterior conditional. This piecewise linear approximation requires a number of discrete sampling points to be specified. Because when we start our model exploration, we know little about the location of the corresponding posterior densities, we used during these first 2,000 MC steps, 100 sampling points regularly spaced on the corresponding parameter domain defined by the prior (PRIOR DENSITY). Figure 4A illustrates the energy evolution of the 10 different trials (realizations). The reader not familiar with our notion of energy (Eq. 16) can think of it as an indication of the quality of fit. The lower it is, the better the fit. Indeed, if we were working with a template-matching paradigm, the sum of the squared errors would be proportional to our energy. The striking feature on Fig. 4A is the presence of "meta-stable" states (a state is defined by a configuration and a value for each model parameter). The trials can spend many steps at an almost constant energy level before making discrete downward steps (like the one falling from 5,800 to 4,300 before the 500th step). If we look at the parameters values and configurations generated by the different trials (not shown), the ones that end above 5,000 "miss" the purple cluster on Fig. 2B. The time required to perform 2,000 MC steps with 7 neurons in the model was roughly 9 min, meaning that 1.5 h was required to obtain the data of Fig. 4A. This time (for 2,000 MC steps) grew from 8' for a model with 4 neurons to 10' for a model with 10 neurons.

View larger version (37K): [in this window] [in a new window]   FIG. 4. A: energy evolution during 2,000 steps with 10 different initial random guesses. Best trial is shown in gray. B: follow-up of the energy evolution of the best trial without (gray) and with (black) the replica exchange method (REM). Circles: mean energy computed from the last 5,000 Monte Carlo (MC) steps of 5 runs without REM. Triangles: mean energy computed from the last 1,000 MC steps of 5 runs with REM.

MODEL CHOICE. Figure 5 illustrates what we meant by "empirical features associated with different models" at the beginning of this section. We have now performed the same series of runs with 7 different models having between 4 and 10 neurons. One feature we can look at in the output of these runs is the most likely classification produced (Empirical averages and SDs, Eq. 21). By "most likely" we mean that the label of each spike has been forced to its most likely value. Figure 5A shows one of the Wilson plots with different neuron numbers, where the events have been colored according to their most likely label. The good behavior of the algorithm appears clearly. When asked to account for the data with 6 neurons, it mainly lumps together the "small" neuron (yellow on Fig. 2B, neuron 6 in Table 1) with the "noise" neuron (black on Fig. 2B, neuron 7 in Table 1). With 7 neurons, the actual number of neurons, it splits the previously lumped clusters into 2 (although it attributes too many spikes to the small neuron and too few to the noise one; see Long runs with full data set). With 8 neurons, it further splits the noise cluster. In fact it creates a "noise" cluster (black on the figure) whose peak amplitude on the fourth recording site is roughly 3 times the recording noise SD and another one (clear blue) whose peak amplitude on this same site is smaller than 3. We remind the reader that our detection threshold is here set at 3 noise SD. The algorithm does not split the doublet-firing neuron (red) into 2 neurons, although this neuron gives rise to two well-separated clusters. With 9 neurons, it splits even further the previous clear blue noise "neuron" into 2 neurons, one with a peak amplitude roughly equal to the detection threshold on the first recording site (pink cluster) and one whose peak amplitudes on sites 1 and 4 are smaller than 3. With 10 neurons (not shown) we end up with the actual noise neuron, resulting in 4 neurons with an average peak amplitude slightly above the detection threshold on one of the recording sites and very close to zero on the 3 others.

View larger version (19K): [in this window] [in a new window]   FIG. 5. Model choice. A: most likely configurations generated by 4 of the 7 models studied are displayed on a Wilson plot where the amplitude on site 4 is plotted against the amplitude on site 1. Scale bars as in Fig. 2. B: maximal peak amplitude plots showing that neuron identification from model to model is easy. Each semiaxis corresponds to the maximal peak amplitude (estimated from the algorithm output) for each of the 6 "important neurons." Vertical axis in the upper half-plan corresponds to the maximal peak amplitude on site 1. Horizontal axis in the right half-plan corresponds to the maximal peak amplitude on site 2. Vertical axis in the lower half-plan corresponds to the amplitude on site 3 and the horizontal axis in the left half-plan corresponds to the amplitude on site 4. For each semiaxis the amplitude value goes from 0 at the origin to 20 at the tip. Each neuron in each model corresponds to a frame. To make the figure more readable (by avoiding a too strong overlap of the frames) a random Gaussian noise (µ = 0, = 0.1) has been added to the maximal peak amplitude values of each neuron.

View larger version (30K): [in this window] [in a new window]   FIG. 7. Marginal posterior density estimate for each parameter of the model computed from the last 2,000 MC steps performed.

The analysis of the full data set (5,058 spikes or 15 s of data) was done in 2 stages. During the first 4,000 MC steps we fixed the model parameters at their mean values computed from the last 1,000 MC steps of the REM run with the reduced data set (Fig. 4B, black trace) and we updated only the configuration. The initial configuration was, as usual, randomly set. We used the last 1,000 MC steps of this first stage to take "measurements" and compare the algorithm's output with the actual values of different data statistics (Figs. 8 and 9). The "complete" algorithm was used during the 21,000 MC steps of the second stage. By complete algorithm we mean that at each MC step a new configuration was drawn as well as new values for the model parameters. The 13 sampling points of the posterior conditionals for the amplitude parameters between steps 4,001 and 6,000 were set using the last 1,000 MC steps of the REM run with the reduced data set. The last 2,000 MC steps of the second stage were used to take measurements. We use here a very long run to illustrate the behavior of our algorithm. Such long, and time-consuming, runs are not required for a daily use of our method. The time required to run the first stage (4,000 MC steps) was 3 h, whereas 26 h were required for the second (21,000 MC steps).

View larger version (17K): [in this window] [in a new window]   FIG. 8. Comparison between the actual configuration and the most likely configuration generated by the algorithm. Two measurements periods are considered, the first one between steps 3,001 and 4,000 (Fig. 6A) represented here in black and the second period (from step 23001 to step 25000) represented in gray. A: number of false negative for each neuron. B: number of false positive for each neuron. C: fraction of wrongly classified spikes for each neuron.

View larger version (48K): [in this window] [in a new window]   FIG. 6. A: energy evolution during a long run with a full data set (5,058 events) and a model with 7 neurons. Fifteen inverse temperatures used were: 1, 0.9, 0.8, 0.7, 0.6, 0.55, 0.5, 0.475, 0.45, 0.425, 0.4, 0.375, 0.35, 0.325, and 0.3. During the first 4,000 steps the algorithm updated only the configuration, not the parameters that were fixed at their average values obtained during the runs with the reduced data set. Black horizontal bar indicates the first measuring period (between steps 3,001 and 4,000), the gray one, the second measuring period (between steps 23,001 and 25,000). B: energy histograms obtained from the second measuring period. Left histogram corresponds to = 1, the right one to = 0.3. C: energy autocorrelation functions from the second measuring period. D: path of the first replica.

A more compact way to present the information of Fig. 7 is to summarize the distributions that are approximately Gaussian by their means and autocorrelation corrected SDs. The other distributions can be summarized by a 95% confidence interval, the left boundary being such that 2.5% of the generated parameters values were below it and the right boundary such that 2.5% of the generated values were above it, as shown in Table 2 (compare with the actual values of Table 1).

ISI HISTOGRAMS ESTIMATES. The last statistics shown are the ISI histograms. Our algorithm generates at each MC step a new configuration, that is, a new set of labels for the spikes. An ISI histogram can therefore be computed for each neuron after each algorithm step. Then the best estimate we can get for an actual histogram (i.e., the histogram obtained using the true labels) is the average of the histograms computed after each step. This is better than getting first the most likely configuration (as we did in the previous paragraph) and then the histogram for each neuron because this latter method introduces a bias in the estimate. We have done this computation for each neuron during the 2 periods. The results together with the actual histograms are shown on Fig. 9. Again, our algorithm generates very good estimates for the 5 good neurons. In particular it generates the proper strongly bimodal ISI histogram for the doublet-firing neuron (Fig. 9C), whereas the model does use a unimodal log-Normal ISI density to describe this neuron. Here again that gain in accuracy provided by the very long run with the complete algorithm is rather modest.

	DISCUSSION

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
We have described an application of the MCMC methodology to the spike-sorting problem. This approach was recently applied to the same problem (Nguyen et al. 2003) but with a simpler data generation model, which led to a simpler algorithm because successive spikes were considered as independent. We have shown here that a MCMC algorithm allows the neurophysiologist to use much richer data generation models than he/she could previously afford, thereby allowing him/her to include a description of the ISI density as well as a description of the spike waveform dynamics. For the sake of illustration we have used in this paper simple but nontrivial models of ISI densities and waveform dynamics. It should nevertheless be clear that the MCMC approach is very flexible (perhaps too flexible). It is for instance very easy to include more sophisticated models for the neuronal discharge based on hidden Markov chains, like the one underlying the doublet-firing neuron of the present paper (Delescluse et al. unpublished observations). We have shown as well for the first time as far as we know, how to estimate both a probability density for the spike train configuration (i.e., the soft classification) and a probability density for the model parameters. The way to use the soft classification to compute statistics of interest (e.g., ISI histograms) was moreover illustrated.

We have shown the presence of meta-stable states, which are, we think, an intrinsic property of models with "interacting" spikes. We have illustrated the potential problems resulting from these meta-stable states, which could introduce a bias in our algorithm's output. However, using a powerful methodology recently introduced in statistics and computational physics, the replica exchange method, we were able to circumvent the slow relaxation resulting from the meta-stable states.

We have deliberately used simulated data that did not correspond to our model assumptions, but which were hopefully realistic enough. Our algorithm with its "simple" data generation model performed very well for the neurons that generated events larger than the detection threshold, even for a neuron which generated two separated clusters on the Wilson plots. As far as we know, no other algorithm would automatically properly cluster the events of such a neuron. Clearly, such a performance is possible only if both the spike occurrence times and amplitude dynamics are taken into account. The robustness of the algorithm's output with respect to overclustering was moreover demonstrated. That being said, if a recording noise model based on a Student's t distribution turns out to be better than a Gaussian model, our algorithm can be trivially modified to work with the former. It can be easily modified as well to include an explicit "noise neuron" (i.e., a neuron with an exponential ISI density and a maximal amplitude below the detection threshold on all recording sites). We permit reader to download our implementation of the algorithm and check that an almost perfect classification (and model parameter estimation) can be obtained for data that correspond to the model, even when strong overlap between clusters are observed on the Wilson plots.

We have considered spikes described by their peak amplitude(s) instead of their waveforms, although it would not require a tremendous algorithmic change to deal with waveforms (see for instance Pouzat et al. 2002). Caution would nevertheless be required to deal with superpositions of spikes, but as long as spikes do not occur exactly at the same time, superpositions will show up as "interaction" terms. That is, Eq. 2 would have to be modified to include a baseline change because of neighboring spikes.

The major drawback of our algorithm is the rather long computational time it requires. It should nevertheless be clear that the user can choose between different implementations of our approach. A fast one, where the model parameters are set during a prerun with a reduced data set, leads to an already very accurate classification of spikes from "good" neurons (a good neuron generates events clearly above the detection threshold). We think that this use of our method should satisfy most practitioners. For a high accuracy of the model parameters estimates the "full" version of our algorithm should be chosen. This version is now clearly too time consuming to be systematically used. This long computational time can be easily reduced, however, if one realizes that the REM can be trivially parallelized (Hansmann 1997; Hukushima and Nemoto 1996; Mitsutake et al. 2001). That is, if one has 15 PCs (which is not such a big investment anymore), one can simulate each replica of Long runs with full data set on a single CPU and when the replica exchange is accepted just "swap the inverse temperatures of the different CPUs" (rather than swapping the states of replicas simulated on different CPUs). We can therefore expect to reduce the computation time by an order of magnitude (even more if one considers that recent CPUs run at 3 GHz, whereas the one used in this paper ran at 1.6 GHz). Preliminary tests on a Beowulf cluster confirmed this expectation. Other improvements can be brought as well, in particular for the amplitude parameters generation. For the latter, a Langevin diffusion or a simple random walk can be used to generate the proposed moves (Besag 2001; Celeux et al. 2000; Liu 2001; Neal 1993). We have in fact already done it (Pouzat et al. unpublished observations) and found that the computation time required for the amplitude parameters was reduced by a factor of 10 (compared with a piecewise linear approximation of the posterior conditional based on 13 sample points).

Finally, as we mentioned in the INTRODUCTION, our method is semiautomatic because user input is still required to choose the "correct" number of active neurons in the data. To make the method fully automatic we will have to do better and that fundamentally means estimating the normalizing constant Z of Eqs. 12 and 13, which is the probability of the data. Indeed Bayesian model comparison requires the comparison (ratio) of data probability under different models (Gelman and Meng 1998; Green 1995) as explained, in a spike-sorting context, by Nguyen et al. (2003). This task of estimating Z could seem daunting but luckily is not. As we said, Dynamic MC and MCMC methods have been used for a long time, which means that other people in other fields already had that problem. Among the proposed solutions (Gelman and Meng 1998; Green 1995; Nguyen et al. 2003), what physicists call thermo-dynamic integration (Frenkel and Smit 2002; Landau and Binder 2000) seems very attractive because it is known to be robust and it requires simulations to be performed between = 1 and = 0. That is precisely what we are already (partially) doing with the REM. Of course the estimation of Z is only half of the problem. The prior distribution on the number of neurons in the model has to be set properly as well. Increasing the number of neurons will always lead to a decrease in energy which will give larger Z values (the derivative of the log of Z being proportional to the opposite of the energy; Frenkel and Smit 2002), we will therefore have to compensate for this systematic Z increase with the prior distribution (Pouzat et al. unpublished observations).

	APPENDIX

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
The Metropolis–Hastings algorithm

The theory of Markov chains (Brémaud 1998) tells us that given a probability density like _post, and a transition kernel T, an equation like Eq. 20 will hold for any function E and for any initial "state" [⁽⁰⁾, C⁽⁰⁾], if T is irreducible and aperiodic and if _post is stationary with respect to T, that is, if

(A1)

In words, irreducible means that the Markov chain obtained by applying T repetitively can reach any "state" (, C) in a finite number of steps. Aperiodic means, loosely speaking, that there is no temporal order in the way the chain returns to any state it visited. These 2 properties are often lumped together in the notion of ergodicity. In practice, if a transition kernel is nonzero for any pair of arbitrary states (_a, C_a) and (_b, C_b), then it is both irreducible and aperiodic (ergodic). Our problem thus becomes to find an ergodic T for which Eq. A1 holds.

It turns out that there is a very general prescription to build a transition kernel T with the desired properties and that, in fact, we do not even have to build it explicitly. We first need a "proposal" transition kernel g[, C|^(t-1), C^(t-1)], which is itself ergodic on the same space as _post.

We then apply the following algorithm:

Given a state [^(t-1), C^(t-1)]

Generate:

(A2)

Take:

where the acceptance probability A is defined by

(A3)

It is not hard to show that the transition kernel induced by this algorithm has the desired properties (Liu 2001; Robert and Casella 1999). The important feature of this procedure is that its implementation requires only a knowledge of _post up to a normalizing constant because Eq. A3 involves a ratio of _post values in 2 states. In other words it requires only the capability to compute the energy of any state. An initial version of this algorithm was given by Metropolis et al. (1953) and the present version is attributed to Hastings (1970).

The reader can see that the Metropolis–Hastings (MH) algorithm is indeed a very large family of algorithms because there exists a priori an infinite number of transitions g from which a specific procedure can be built. We therefore have to find at least one g and if we find several we would like to have a way to compare the resulting MH transition kernels (T). A common way to proceed for complicated models, like the one we are presently dealing with, is to split the transition kernel T in a series of parameter and label specific transition kernels (Besag 2001; Fishman 1996). Let us define

(A4)

where n_p is the number of parameters in the model and

(A5)

Then the parameter-specific transition kernels are objects like

(A6)

and the label specific transition kernels

(A7)

We will often use the following short notation for Eqs. A6 and A7

(A8)

and

(A9)

A "complete" transition kernel can be constructed with, for instance, a sequential application of every label and parameter specific transition

(A10)

In the sequel we will say that one MC time step has been performed every time a complete transition kernel has been applied to the "present" state of the Markov chain. Sufficient conditions the parameter and label specific transitions must exhibit for the complete transition T to be ergodic and for _post to be stationary with respect to T are as follows:

1. Each parameter and label specific transition (T_-i, T_C-j) is such that
   

   (A11)

   for all pairs of states (_-i, _i = a, C) and (_-i, _i = b, C) such that
   

   (A12)

   (It is straightforward to write the equivalent condition for T_C-j)
   

2. Each parameter and label specific transition (T_-i, T_C-j) is such that the detailed balance condition is met
   

   (A13)

   Moreover, we have
   

   and Eq. A12 implies that
   

   We can therefore rewrite the detailed balance condition as follows
   

   (A14)

   where _post(_i|_-i, C, Y) is the posterior conditional density of _i. There are, of course, equivalent equations for the label-specific transition kernels (T_C).
   

A straightforward way to obtain T_-i and T_C-i with the desired properties is to find parameter- and label-specific proposal transition kernels (g_-i and g_C-i) respecting Eq. A11 in combination with an acceptance probability like Eq. A3, that is, to have an MH rule for each parameter of the model and for each label of the configuration. It is indeed easy to show that such a construct gives rise to T_-i and T_C-i respecting the detailed balance condition because, assuming a b

(49)

To complete this justification of this parameter and label specific decomposition the reader needs to realize that the detailed balance condition is stronger than the stationarity condition (Eq. A1) and in fact implies it as shown in the following paragraph.

The detailed balance condition imposed on the parameter specific transition kernels implies the stationarity condition on the resulting "complete" transition kernel.We present here the justification for a simple model with 2 parameters (₁ and ₂) and without latent variable, the extension to the full model considered in this paper being straightforward. So we have

with the detailed balance conditions:

(A15)

and

(A16)

We then have

(A17)

(A18)

(A19)

(A20)

(A21)

(A22)

(A23)

where the detailed balance conditions (Eqs. A15 and A16) have been used to go from Eq. A17 to Eq. A18 and from Eq. A20 to Eq. A21, respectively and the normalization condition for a proper transition kernel

(60)

has been used to go from Eq. A19 to Eq. A20 and from Eq. A22 to Eq. A23.

It should become clear in the sequel that the appeal of the detailed balance comes from the (relative) ease with which it can be implemented (and checked). This is not the case of the stationarity condition which is much harder to check for an arbitrary transition kernel.

A close examination of the MH acceptance probability shows that a reasonable choice for the parameter specific proposal transition would be the corresponding conditional posterior density. For then we would have

That is, with the posterior conditional as a proposal transition kernel the acceptance probability is always 1. This type of parameter specific transition is called heat-bath algorithm by physicists (Landau and Binder 2000; Newman and Barkema 1999; Sokal 1989) and Gibbs sampler by statisticians (Gelman et al. 1995; Liu 2001; Robert and Casella 1999). It has already been used in a neurophysiological context, for single-channel studies (Ball et al. 1999; Rosales et al. 2001) and even for spike-sorting (Nguyen et al. 2003). It is in general a good first trial when one wants to build a Markov chain on a parameter and configuration space like the one we are considering here. Its main drawback is that it requires the expression of the posterior conditional to be available in a closed form. For the present model, such a closed form posterior conditional is available for the labels and for the scale and shape parameters of the ISI density, but not for the "amplitude parameters" (P, , ). The following section of the APPENDIX contains a detailed account of the algorithms used for the labels transition kernels (SPIKE LABEL TRANSITION MATRIX), the scale parameter transition kernel (SCALE PARAMETER TRANSITION KERNEL), the shape parameter transition kernel (SHAPE PARAMETER TRANSITION KERNEL), and the amplitude parameters transition kernels (AMPLITUDE PARAMETERS TRANSITION KERNELS). The algorithm of the later uses as proposal transition kernels, piecewise linear approximations of the posterior conditional densities. These piecewise linear approximations are computationally costly (they take a lot of time to compute) when one uses a lot of discrete posterior conditional evaluations, although they are then better approximations. To keep both a good precision and a reasonable computational cost we used a multiruns procedure where the sample generated during one run was used to optimize the few (13) locations at which the posterior conditional was evaluated (AMPLITUDE PARAMETERS TRANSITION KERNELS). This "position refinement" procedure is one of the reasons why we repetitively use successions of runs in the results section.

Parameter-specific transition kernels

SPIKE LABEL TRANSITION MATRIX. Here c_i takes integer values in {1,..., K}, so we are therefore, strictly speaking, dealing with a transition matrix rather than a transition kernel. As explained in The Metropolis–Hastings algorithm we build T_C-i as a product of a proposal transition matrix g_C-i and an acceptance probability "matrix" A_C-i. We always propose a label different from the present one [g_C-i (c_i = a|c_i = a) = 0], and the "new" labels are proposed according to their posterior conditional density

(A24)

The acceptance probability is defined by Eq. A3 and gives here

(A25)

We will illustrate the computation on the simple case presented in LIKELIHOOD FUNCTION and Fig. 1, with one recording site and 2 neurons in the model (K = 2). Because c_i can only have 2 values, c_i = 1 and c_i = 2, the matrix g_C-i is straightforward to write

The acceptance probability requires the calculation of the ratio of the posterior conditionals: _post(c_i = 1|, C_-i, Y) and _post(c_i = 2|, C_-i, Y). From Eq. 15 we have

Therefore

If we then assume for instance that

we get the following transition matrix

In the case of Fig. 1, if we consider i = 13, we get, using Eq. 5

and

where M represents all the terms that are identical in the 2 equations.

A remark on numerical implementation.To avoid rounding errors and keep numerical precision we work with the log-likelihoods

look for l_max, the maximum (with respect to b)of l_b and obtain _post(c_i = b|, C_-i, Y) with

A proposed value b {1,..., K}\{a} is drawn from the probability mass g_C-i (c_i = b|c_i = a) using the inverse cumulative distribution function (Fishman 1996; Gelman et al. 1995; Robert and Casella 1999).

Boundary treatment. From the statisticians' viewpoint the spike trains we are dealing with are censored data. They mean by that that there are missing data without which we cannot, strictly speaking, compute our likelihood function. The missing data here are the last spikes from each neuron that occurred before our recording started and the next spikes after our recording stopped. There are proper approaches to deal with such censored data (Gelman et al. 1995; Robert and Casella 1999) and we plan to implement them in a near future. In the present study we used a simple and quick way to deal with the end effects in our data. We used periodic boundary conditions; that is, the last spike from each neuron was considered as the spike preceding the first spike of the same neuron.

SCALE PARAMETER TRANSITION KERNEL. We consider here cases where _i of Eq. A14 corresponds to one of the K scale parameters, s_q, where q stands for the neuron index. We will first show that the corresponding posterior conditional: _post(_i = s_q|_-i, C, Y) can be written in a closed form. From Eq. 15 we have

Going back to the structure of the likelihood function (Eq. 9), one sees, using Eqs. 5 and 10, that the likelihood function with the labels set and all model parameters set except the scale parameter of one neuron q depends only on the events that are labeled as belonging to neuron q, and on the shape parameter of q, _q. If we write {i_q,1,..., i_q,n}, the ISI computed from the spike train of neuron q in configuration C, we get

If we introduce: , a short manipulation shows that

In words: the posterior density of the log of the scale parameter s_q is Normal with mean: , and variance, (_q)²/n_q. Our prior on s_q is a uniform distribution between s_min and s_max (PRIOR DENSITY); we can therefore, using the Gibbs algorithm, generate s_q as follows:

1. Given C and Y: compute n_q and
   
2. Generate u Norm
   
3. Compute s = exp(u)
   
4. If s [s_min, s_max], set s_q = s; otherwise, go back to 2
   

SHAPE PARAMETER TRANSITION KERNEL. Here _i is the shape parameter of neuron q, _i = _q. We again have to show that the posterior conditional can be written in a closed form. Using the approach of the previous section we quickly get

Then an identification with an inverse-Gamma density gives us the answer

if we use = (_q)², = (n_q/2) - 1, and = n_q/2( - log s_q)². To generate _q with the Gibbs algorithm we need to know how to generate random number with an inverse-Gamma density. It turns out that most analysis platforms (Scilab, MATLAB) and C libraries like GSL do not have an inverse-Gamma random number generator but do have a Gamma random number generator. A Gamma density can be written

and it is not hard to show that if is Gamma with parameters and , then 1/ is inverse-Gamma with the same parameters. To generate _q using the Gibbs algorithm, we therefore perform

1. Generate
   
2. if , then otherwise go back to 1.
   

AMPLITUDE PARAMETERS TRANSITION KERNELS. As explained in The Metropolis–Hastings algorithm the amplitude parameters: P, , cannot be generated with the Gibbs algorithm because the corresponding posterior conditionals cannot be written in a closed form. For these parameters we had recourse to a piecewise linear proposal kernel. The principle of our method is illustrated here with a simple example, which can be thought of as the generation of the maximal peak amplitude of one of the neurons (q) on, say, the first recording site. Using a discrete set of points on the domain allowed by the corresponding priors: {P_q,1,1 = P_min, P_q,1,2,..., P_q,1,m = P_max}, an unnormalized value of the posterior conditional is computed

Again the structure of the likelihood (Eq. 9) allows us to write

where we are assuming we are dealing with a tetrode recording. These unnormalized posterior values are then normalized such that the piecewise linear function:

where P [P_q,1,i, P_q,1,i+1] sums to 1. Of course when we start the algorithm we do not know what is the best location for the discrete points P_q,1,i. We therefore used during a first run 100 regularly spaced points, although it turns out that computing at each MC step 100 values of the posterior conditional (for each amplitude parameter of the model) takes time. It is therefore very interesting, to speed up the sorting procedure, to use fewer points, which basically means forgetting about the points located where the posterior conditional is extremely small. To achieve this goal, at the end of each run, the last 1000 generated values of the parameter of interest (P_q,1) were used to get an estimate of the posterior marginal cumulative distribution of the parameter. Then 10 quantiles were located on this cumulative distribution, one discrete point was taken slightly smaller than the smallest generated value for that parameter, and the two extreme values allowed by the priors were used as discrete points as well. The new piecewise proposal kernel was therefore based on a sparser approximation.

Slow relaxation and REM

One of the nice features of the MCMC approach, and in fact the key theoretical result justifying its growing popularity, is that the user can be sure that empirical averages (Eq. 19) computed from the samples it generates do converge to the corresponding expected values (Eq. 18). The not-so-nice feature, forgetting about the actual implementation details, which can be tedious, is that the user can run his/her MCMC algorithm only for a finite amount of time. Then the fact that empirical averages are approximations cannot be ignored. There are 3 sources of errors in such estimates. The first that comes to mind is the one resulting from the finite sample size, that is, the one we would have even if our draws where directly generated from _post. The difference between the MCMC-based estimates and estimates based on direct samples is that the variance of the estimators of the former have to be corrected to take into account the correlation of the states generated by the Markov chain, as explained in Empirical averages and SDs. The second source of error is a bias induced by the state [⁽⁰⁾, C⁽⁰⁾] with which the chain is initialized (Fishman 1996; Sokal 1989). The bad news concerning this source of error is that there is no general theoretical result providing guidance on the way to handle it, but the booming activity in the Markov chain field already produced encouraging results in particular cases (Liu 2001). The common wisdom in the field is to monitor parameters (and labels) evolution, and/or their functions like the energy (Eq. 16). Based on examination of evolution plots (e.g., Fig. 4, A and B, Fig. 6A) and/or on application of time-series analysis tools, the user will decide that "equilibrium" has been reached and discard the parameters values before equilibrium. More sophisticated tests do exist (Robert and Casella 1999) but they were not used in this paper. These first 2 sources of error are common to all MCMC approaches. The third one appears only when the energy function exhibits several local minima. In the latter case, the Markov chain realization can get trapped in a local minimum, which could be a poor representation of the whole energy function. This sensitivity to local minima arises from the local nature of the transitions generated by the MH algorithm. That is, in our case, at each MC time step, we first attempt to change the label of spike 1, then the one of spike 2, and so forth, then the one of spike N, then we try to change the first component of (P_1,1), and so on until the last component (_K). That implies that if we start in a local minimum and if we need to change, say, the labels of 10 spikes to reach another lower local minimum, we could have a situation in which the first 3 label changes are energetically unfavorable (giving, for instance, an acceptance probability, Eq. A3, of 0.1 per label change), which would make the probability to accept the succession of changes very low (0.1³), meaning that our Markov chain would spend a long time in the initial local minimum before "escaping" to the neighboring one. Stated more quantitatively, the average time the chain will take to escape from a local minimum with energy E_min grows as the exponential of the energy difference between the energy E* of the highest energy state the chain has to go through to escape and E_min

Our chains will therefore exhibit an Arrhenius behavior. To sample more efficiently such state spaces, the replica exchange method (REM) (Hukushima and Nemoto 1996; Mitsutake et al. 2001), also known as the parallel tempering method (Falcioni and Deem 1999; Hansmann 1997; Yan and de Pablo 1999), considers R replicas of the system with an increasing sequence of temperatures (or a decreasing sequence of ) and a dynamic defined by 2 types of transitions: usual MH transitions performed independently on each replica according to the rule defined by Eq. A10 and a replica exchange transition. The key idea is that the high temperature (low ) replicas will be able to easily cross the energy barriers separating local minima (in the example above, if we had a probability of 0.1³ to accept a sequence of labels switch for the replica at = 1, the replica at = 0.2 will have a probability 0.1^3x0.2 0.25 to accept the same sequence). What is needed is a way to generate replica exchange transitions such that the replica at = 1 generates a sample from _post defined by Eq. 17. Formally the REM consists in simulating, on an "extended ensemble," a Markov chain whose unique stationary density is given by

(A26)

where "ee" in _ee stands for "extended ensemble" (Iba 2001), R is the number of simulated replicas, ₁ >... > _R for convenience, and

(A27)

That is, compared to Eq. 17, we now explicitly allow to be different from 1. To construct our "complete" transition kernel we apply our previous procedure; that is, we construct it as a sequence of parameter, label, and interreplica-specific MH transitions. We already know how to obtain the parameter and label specific transitions for each replica. What we really need is a transition to exchange replicas, say the replicas at inverse temperature _i and _i+1, such that the detailed balance is preserved (Eq. A13)

which leads to

Again we write T_i,i+1 as a product of a proposal transition kernel and an acceptance probability. Here we have already explicitly chosen a deterministic proposal (we only propose transitions between replicas at neighboring inverse temperatures), which gives us

It is therefore enough to take

(A28)

The reader sees that if the state of the "hot" replica (_i+1) has a lower energy [E(_i, C_i)] than the "cold" one, the proposed exchange is always accepted. The exchange can pictorially be seen as cooling down the hot replica and warming up the cold one. Fundamentally this process amounts to making the replica that is at the beginning and at the end of the replica exchange transition at the cold temperature to jump from one local minimum (_i+1, C_i+1) to another one (_i, C_i). That is precisely what we were looking for. The fact that we can as well accept unfavorable exchanges (i.e., raising the energy of the "cold" replica and decreasing the one of the "hot" replica) is the price we have to pay for our algorithm to generate samples from the proper posterior (we are not doing optimization here).

For the replica exchange transition to work well we need to be careful with our choice of inverse temperatures. The typical energy of a replica (i.e., its expected energy) increases when decreases (Fig. 6A). We will therefore typically have a positive energy difference: E = E_hot - E_cold > 0 between the replicas at low and high before the exchange. That implies that the acceptance ratio (Eq. A28) for the replica exchange will be typically smaller than 1. Obviously, if it becomes too small, exchanges will practically never be accepted. To avoid this situation we need to choose our inverse temperatures such that the typical product: E, where = _cold - _hot, is close enough to zero (Hukushima and Nemoto 1996; Iba 2001; Neal 1994). In practice we used preruns with an a priori too large number of s, checked the resulting energy histograms and kept enough inverse temperatures to have some overlap between successive histograms (Fig. 6B).

The replica exchange transitions were performed between each pair _i, _i+1 with an even, respectively odd, i at the end of each even, respectively odd, MC time step. With the replica exchange scheme, each MC time step was therefore composed of a complete parameter and label transition for each replica, followed by a replica exchange transition. This scheme corresponds to the one described by Hukushima and Nemoto (1996). A rather detailed account of the REM can be found in Mitsutake et al. (2001). Variations on this scheme do exist (Celeux et al. 2000; Neal 1994).

Random number generators for the simulated data

A vector n from an isotropic t density with (>2) degrees of freedom and a SD equal to 1 (for each component) was generated with the following algorithm (Fishman 1996; Gelman et al. 1995):

1. Generate: z Norm (0, 1)
   
2. Generate: x ²()
   
3. Take:
   

The Normal and ² (pseudo-)random number generators are built-in features of Scilab (and MATLAB).

An isi from a log-Normal density with scale parameter s and shape parameter was generated with the following algorithm:

1. Generate: z Norm (0, ²)
   
2. Take: isi = s exp(z)
   

For the neuron with a Gamma ISI density, the Gamma random number generator of Scilab was used.

	ACKNOWLEDGMENTS

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
We thank C. van Vreeswijk, N. Brunel, O. Mazor, L. Forti, A. Marty, J. Chavas, and M.-G. Lucas for comments and suggestions on the manuscript.

GRANTS

This work was supported in part by a grant from the Ministère de la Recherche (ACI Neurosciences Intégratives et Computationnelles, pré-projet, 2001–2003), by a grant from inter–Établissements Publiques Scientifiques et Techniques (Bioinformatique), and by the région Ile-de-France (programme Sésame). M. Delescluse was supported by a fellowship from the Ministère de l'Education National et de la Recherche.

	FOOTNOTES

 
The costs of publication of this article were defrayed in part by the payment of page charges. The article must therefore be hereby marked "advertisement" in accordance with 18 U.S.C. Section 1734 solely to indicate this fact.

Address for reprint requests and other correspondence: C. Pouzat, Laboratoire de Physiologie Cérébrale, CNRS UMR 8118, Université René Descartes, 45 rue des Saints Pères, 75006, Paris, France (E-mail: christophe.pouzat{at}univ-paris5.fr).

	REFERENCES

TOP ABSTRACT INTRODUCTION METHODS RESULTS DISCUSSION APPENDIX ACKNOWLEDGMENTS REFERENCES

 
Ball FG, Cai Y, Kadane JB, and O'Hagan A. Bayesian inference for ion-channel gating mechanisms directly from single-channel recordings, using Markov chain Monte Carlo. Proc R Soc Lond A Phys Sci 455: 2879-2932, 1999.[Abstract/Free Full Text]

Besag J. Markov Chain Monte Carlo for Statistical Inference [online]. Working Paper No. 9. Seattle, WA: Center for Statistics and Social Sciences, Univ. of Washington, 2001. http://www.csss.washington.edu/Papers/wp9.pdf [28 Nov. 2003].

Binder K and Young AP. Spin glasses: experimental facts, theoretical concepts and open questions. Rev Mod Phys 58: 801-976, 1986.[CrossRef][Web of Science]

Brémaud P. Markov Chains: Gibbs Fields, Monte Carlo Simulations and Queues. New York: Springer-Verlag, 1998.

Celeux G, Hurn M, and Robert C. Computational and inferential difficulties with mixture posterior distributions. J Am Stat Assoc 95: 957-970, 2000.[CrossRef][Web of Science]

Falcioni M and Deem MW. A biased Monte Carlo scheme for zeolite structure solution. J Chem Phys 110: 1754-1766, 1999.[CrossRef]

Fee MS, Mitra PP, and Kleinfeld D. Automatic sorting of multiple-unit neuronal signals in the presence of anisotropic and non-Gaussian variability. J Neurosci Methods 69: 175-188, 1996a.[CrossRef][Web of Science][Medline]

Fee MS, Mitra PP, and Kleinfeld D. Variability of extracellular spike waveforms of cortical neurons. J Neurophysiol 76: 3823-3833, 1996b.[Abstract/Free Full Text]

Fishman GS. Monte Carlo, Concepts, Algorithms and Applications. New York: Springer-Verlag, 1996.

Frenkel D and Smit B. Understanding Molecular Simulation. From Algorithms to Applications. San Diego, CA: Academic Press, 2002.

Gelman A, Carlin JB, Stern HS, and Rubin DB. Bayesian Data Analysis. Boca Raton, FL: Chapman & Hall/CRC, 1995.

Gelman A and Meng XL. Simulating normalizing constants: from importance sampling to bridge sampling to path sampling. Stat Sci 13: 163-185, 1998.[CrossRef][Web of Science]

Geman S and Geman D. Stochastic relaxation, Gibbs distributions and the Bayesian restoration of images. IEEE Trans Pattern Anal Mach Intell 6: 721-741, 1984.[CrossRef]

Gray CM, Maldonado PE, Wilson M, and McNaughton B. Tetrodes markedly improve the reliability and yield of multiple single-unit isolation from multi-unit recordings in cat striate cortex. J Neurosci Methods 63: 43-54, 1995.[CrossRef][Web of Science][Medline]

Green PJ. Reversible jump Markov chain Monte Carlo computation and Bayesian model determination. Biometrika 82: 711-732, 1995.[Abstract/Free Full Text]

Hansmann UHE. Parallel tempering algorithm for conformational studies of biological molecules. Chem Phys Lett 281: 140-150, 1997.[CrossRef][Web of Science]

Harris KD, Henze DA, Csicsvari J, Hirase H, and Buzsaki G. Accuracy of tetrode spike separation as determined by simultaneous intracellular and extracellular measurements. J Neurophysiol 84: 401-414, 2000.[Abstract/Free Full Text]

Hastings WK. Monte Carlo sampling methods using Markov chains and their applications. Biometrika 57: 92-109, 1970.

Hukushima K and Nemoto K. Exchange Monte Carlo and application to spin glass simulations. J Phys Soc Jpn 65: 1604-1608, 1996.[CrossRef]

Iba Y. Extended ensemble Monte Carlo. Int J Mod Phys C 12: 623-656, 2001.[CrossRef]

Janke W. Statistical Analysis of Simulations: Data Correlations and Error Estimation [online]. 2002. http://www.fz-juelich.de/nic-series/volume10 [28 Nov. 2003].

Johnson DH. Point process models of single-neuron discharges. J Comput Neurosci 3: 275-299, 1996.[CrossRef][Web of Science][Medline]

Johnson DH, Tsuchitani C, Linebarger DA, and Johnson MJ. Application of a point process model to responses of cat lateral superior olive units to ipsilateral tones. Hearing Res 21: 135-159, 1986.[CrossRef][Web of Science][Medline]

Landau D and Binder K. A Guide to Monte Carlo Simulations in Statistical Physics. Cambridge, UK: Cambridge Univ. Press, 2000.

Lewicki MS. Bayesian modeling and classification of neural signals. Neural Comput 6: 1005-1030, 1994.[CrossRef][Web of Science]

Lewicki MS. A review of methods for spike-sorting: the detection and classification of neural action potentials. Network Comput Neural Syst 9: R53-R78, 1998.[CrossRef]

Liu JS. Monte Carlo Strategies in Scientific Computing. New York: Springer-Verlag, 2001.

Matsumoto M and Nishimura T. Mersenne twister: a 623-dimensionally equidistributed uniform pseudorandom number generator. ACM Trans Model Comput Simul 8: 3-30, 1998.

Metropolis N, Rosenbluth AW, Rosenbluth MN, Teller AH, and Teller E. Equations of state calculations by fast computing machines. J Chem Phys 21: 1087-1092, 1953.[CrossRef]

Mitsutake A, Sugita Y, and Okamoto Y. Generalized-ensemble algorithms for molecular simulations of biopolymers. Biopolymers 60: 96-123, 2001.[CrossRef][Web of Science][Medline]

National Institute of Standards and Technology (NIST). NIST/SEMATECH e-Handbook of Statistical Methods [online]. http://www.itl.nist.gov/div898/handbook/index.htm [28 Nov. 2003].

Neal RM. Probabilistic Inference Using Markov Chain Monte Carlo Methods [online]. Technical Report CRG-TR-93-1. Toronto, Canada: Department of Computer Science, Univ. of Toronto, 1993. http://www.cs.toronto.edu/~radford/papers-online.html [28 Nov. 2003].

Neal RM. Sampling from Multimodal Distributions Using Tempered Transitions [online]. Technical Report No. 9421. Toronto, Canada: Department of Statistics, Univ. of Toronto, 1994. http://www.cs.toronto.edu/~radford/papers-online.html [28 Nov. 2003].

Newman MEJ and Barkema GT. Monte Carlo Methods in Statistical Physics. Oxford, UK: Oxford Univ. Press, 1999.

Nguyen DP, Frank LM, and Brown EN. An application of reversible-jump Markov chain Monte Carlo to spike classification of multi-unit extracellular recordings. Network Comput Neural Syst 14: 61-82, 2003.[CrossRef]

Pouzat C, Mazor O, and Laurent G. Using noise signature to optimize spike-sorting and to assess neuronal classification quality. J Neurosci Methods 122: 43-57, 2002.[CrossRef][Web of Science][Medline]

Quirk MC and Wilson MA. Interaction between spike waveform classification and temporal sequence detection. J Neurosci Methods 94: 41-52, 1999.[CrossRef][Web of Science][Medline]

Robert CP and Casella G. Monte Carlo Statistical Methods. New York: Springer-Verlag, 1999.

Rosales R, Stark A, Fitzgerald WJ, and Hladky SB. Bayesian restoration of ion channel records using hidden Markov models. Biophys J 80: 1088-1103, 2001.[Web of Science][Medline]

Sahani M. Latent Variable Models for Neural Data Analysis (PhD thesis). Pasadena, CA: California Institute of Technology, 2002.

Sokal AD. Monte Carlo Methods in Statistical Mechanics: Foundations and New Algorithms. Cours de Troisième Cycle de la Physique en Suisse Romande [online]. http://citeseer.nj.nec.com/soka196monte.html [28 Nov. 2003].

Wu FY. The Potts model. Rev Mod Phys 54: 235-268, 1982.[CrossRef][Web of Science]

Yan Q and de Pablo JJ. Hyper-parallel tempering Monte Carlo: application to the Lennard-Jones fluid and the restricted primitive model. J Chem Phys 111: 9509-9516, 1999.[CrossRef]

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