The smfsb R package

Introduction

In the previous post I gave a brief introduction to the third edition of my textbook, Stochastic modelling for systems biology. The algorithms described in the book are illustrated by implementations in R. These implementations are collected together in an R package on CRAN called smfsb. This post will provide a brief introduction to the package and its capabilities.

Installation

The package is on CRAN – see the CRAN package page for details. So the simplest way to install it is to enter

install.packages("smfsb")

at the R command prompt. This will install the latest version that is on CRAN. Once installed, the package can be loaded with

library(smfsb)

The package is well-documented, so further information can be obtained with the usual R mechanisms, such as

vignette(package="smfsb")
vignette("smfsb")
help(package="smfsb")
?StepGillespie
example(StepCLE1D)

The version of the package on CRAN is almost certainly what you want. However, the package is developed on R-Forge – see the R-Forge project page for details. So the very latest version of the package can always be installed with

install.packages("smfsb", repos="http://R-Forge.R-project.org")

if you have a reason for wanting it.

A brief tutorial

The vignette gives a quick introduction the the library, which I don’t need to repeat verbatim here. If you are new to the package, I recommend working through that before continuing. Here I’ll concentrate on some of the new features associated with the third edition.

Simulating stochastic kinetic models

Much of the book is concerned with the simulation of stochastic kinetic models using exact and approximate algorithms. Although the primary focus of the text is the application to modelling of intra-cellular processes, the methods are also appropriate for population modelling of ecological and epidemic processes. For example, we can start by simulating a simple susceptible-infectious-recovered (SIR) disease epidemic model.

set.seed(2)
data(spnModels)

stepSIR = StepGillespie(SIR)
plot(simTs(SIR$M, 0, 8, 0.05, stepSIR),
  main="Exact simulation of the SIR model")

Exact simulation of the SIR epidemic model
The focus of the text is stochastic simulation of discrete models, so that is the obvious place to start. But there is also support for continuous deterministic simulation.

plot(simTs(SIR$M, 0, 8, 0.05, StepEulerSPN(SIR)),
  main="Euler simulation of the SIR model")

Euler simulation of the SIR model
My favourite toy population dynamics model is the Lotka-Volterra (LV) model, so I tend to use this frequently as a running example throughout the book. We can simulate this (exactly) as follows.

stepLV = StepGillespie(LV)
plot(simTs(LV$M, 0, 30, 0.2, stepLV),
  main="Exact simulation of the LV model")

Exact simulation of the Lotka-Volterra model

Stochastic reaction-diffusion modelling

The first two editions of the book were almost exclusively concerned with well-mixed systems, where spatial effects are ignorable. One of the main new features of the third edition is the inclusion of a new chapter on spatially extended systems. The focus is on models related to the reaction diffusion master equation (RDME) formulation, rather than individual particle-based simulations. For these models, space is typically divided into a regular grid of voxels, with reactions taking place as normal within each voxel, and additional reaction events included, corresponding to the diffusion of particles to adjacent voxels. So to specify such models, we just need an initial condition, a reaction model, and diffusion coefficients (one for each reacting species). So, we can carry out exact simulation of an RDME model for a 1D spatial domain as follows.

N=20; T=30
x0=matrix(0, nrow=2, ncol=N)
rownames(x0) = c("x1", "x2")
x0[,round(N/2)] = LV$M
stepLV1D = StepGillespie1D(LV, c(0.6, 0.6))
xx = simTs1D(x0, 0, T, 0.2, stepLV1D, verb=TRUE)
image(xx[1,,], main="Prey", xlab="Space", ylab="Time")

Discrete 1D simulation of the LV model

image(xx[2,,], main="Predator", xlab="Space", ylab="Time")

Discrete 1D simulation of the LV model
Exact simulation of discrete stochastic reaction diffusion systems is very expensive (and the reference implementation provided in the package is very inefficient), so we will often use diffusion approximations based on the CLE.

stepLV1DC = StepCLE1D(LV, c(0.6, 0.6))
xx = simTs1D(x0, 0, T, 0.2, stepLV1D)
image(xx[1,,], main="Prey", xlab="Space", ylab="Time")

Spatial CLE simulation of the 1D LV model

image(xx[2,,], main="Predator", xlab="Space", ylab="Time")

Spatial CLE simulation of the 1D LV model
We can think of this algorithm as an explicit numerical integration of the obvious SPDE approximation to the exact model.

The package also includes support for simulation of 2D systems. Again, we can use the Spatial CLE to speed things up.

m=70; n=50; T=10
data(spnModels)
x0=array(0, c(2,m,n))
dimnames(x0)[[1]]=c("x1", "x2")
x0[,round(m/2),round(n/2)] = LV$M
stepLV2D = StepCLE2D(LV, c(0.6,0.6), dt=0.05)
xx = simTs2D(x0, 0, T, 0.5, stepLV2D)
N = dim(xx)[4]
image(xx[1,,,N],main="Prey",xlab="x",ylab="y")

Spatial CLE simulation of the 2D LV model

image(xx[2,,,N],main="Predator",xlab="x",ylab="y")

Spatial CLE simulation of the 2D LV model

Bayesian parameter inference

Although much of the book is concerned with the problem of forward simulation, the final chapters are concerned with the inverse problem of estimating model parameters, such as reaction rate constants, from data. A computational Bayesian approach is adopted, with the main emphasis being placed on “likelihood free” methods, which rely on forward simulation to avoid explicit computation of sample path likelihoods. The second edition included some rudimentary code for a likelihood free particle marginal Metropolis-Hastings (PMMH) particle Markov chain Monte Carlo (pMCMC) algorithm. The third edition includes a more complete and improved implementation, in addition to approximate inference algorithms based on approximate Bayesian computation (ABC).

The key function underpinning the PMMH approach is pfMLLik, which computes an estimate of marginal model log-likelihood using a (bootstrap) particle filter. There is a new implementation of this function with the third edition. There is also a generic implementation of the Metropolis-Hastings algorithm, metropolisHastings, which can be combined with pfMLLik to create a PMMH algorithm. PMMH algorithms are very slow, but a full demo of how to use these functions for parameter inference is included in the package and can be run with

demo(PMCMC)

Simple rejection-based ABC methods are facilitated by the (very simple) function abcRun, which just samples from a prior and then carries out independent simulations in parallel before computing summary statistics. A simple illustration of the use of the function is given below.

data(LVdata)
rprior <- function() { exp(c(runif(1, -3, 3),runif(1,-8,-2),runif(1,-4,2))) }
rmodel <- function(th) { simTs(c(50,100), 0, 30, 2, stepLVc, th) }
sumStats <- identity
ssd = sumStats(LVperfect)
distance <- function(s) {
    diff = s - ssd
    sqrt(sum(diff*diff))
}
rdist <- function(th) { distance(sumStats(rmodel(th))) }
out = abcRun(10000, rprior, rdist)
q=quantile(out$dist, c(0.01, 0.05, 0.1))
print(q)
##       1%       5%      10% 
## 772.5546 845.8879 881.0573
accepted = out$param[out$dist < q[1],]
print(summary(accepted))
##        V1                V2                  V3         
##  Min.   :0.06498   Min.   :0.0004467   Min.   :0.01887  
##  1st Qu.:0.16159   1st Qu.:0.0012598   1st Qu.:0.04122  
##  Median :0.35750   Median :0.0023488   Median :0.14664  
##  Mean   :0.68565   Mean   :0.0046887   Mean   :0.36726  
##  3rd Qu.:0.86708   3rd Qu.:0.0057264   3rd Qu.:0.36870  
##  Max.   :4.76773   Max.   :0.0309364   Max.   :3.79220
print(summary(log(accepted)))
##        V1                V2               V3         
##  Min.   :-2.7337   Min.   :-7.714   Min.   :-3.9702  
##  1st Qu.:-1.8228   1st Qu.:-6.677   1st Qu.:-3.1888  
##  Median :-1.0286   Median :-6.054   Median :-1.9198  
##  Mean   :-0.8906   Mean   :-5.877   Mean   :-1.9649  
##  3rd Qu.:-0.1430   3rd Qu.:-5.163   3rd Qu.:-0.9978  
##  Max.   : 1.5619   Max.   :-3.476   Max.   : 1.3329

Naive rejection-based ABC algorithms are notoriously inefficient, so the library also includes an implementation of a more efficient, sequential version of ABC, often known as ABC-SMC, in the function abcSmc. This function requires specification of a perturbation kernel to “noise up” the particles at each algorithm sweep. Again, the implementation is parallel, using the parallel package to run the required simulations in parallel on multiple cores. A simple illustration of use is given below.

rprior <- function() { c(runif(1, -3, 3), runif(1, -8, -2), runif(1, -4, 2)) }
dprior <- function(x, ...) { dunif(x[1], -3, 3, ...) + 
    dunif(x[2], -8, -2, ...) + dunif(x[3], -4, 2, ...) }
rmodel <- function(th) { simTs(c(50,100), 0, 30, 2, stepLVc, exp(th)) }
rperturb <- function(th){th + rnorm(3, 0, 0.5)}
dperturb <- function(thNew, thOld, ...){sum(dnorm(thNew, thOld, 0.5, ...))}
sumStats <- identity
ssd = sumStats(LVperfect)
distance <- function(s) {
    diff = s - ssd
    sqrt(sum(diff*diff))
}
rdist <- function(th) { distance(sumStats(rmodel(th))) }
out = abcSmc(5000, rprior, dprior, rdist, rperturb,
    dperturb, verb=TRUE, steps=6, factor=5)
## 6 5 4 3 2 1
print(summary(out))
##        V1                V2               V3        
##  Min.   :-2.9961   Min.   :-7.988   Min.   :-3.999  
##  1st Qu.:-1.9001   1st Qu.:-6.786   1st Qu.:-3.428  
##  Median :-1.2571   Median :-6.167   Median :-2.433  
##  Mean   :-1.0789   Mean   :-6.014   Mean   :-2.196  
##  3rd Qu.:-0.2682   3rd Qu.:-5.261   3rd Qu.:-1.161  
##  Max.   : 2.1128   Max.   :-2.925   Max.   : 1.706

We can then plot some results with

hist(out[,1],30,main="log(c1)")

ABC-SMC posterior for the LV model

hist(out[,2],30,main="log(c2)")

ABC-SMC posterior for the LV model

hist(out[,3],30,main="log(c3)")

ABC-SMC posterior for the LV model

Although the inference methods are illustrated in the book in the context of parameter inference for stochastic kinetic models, their implementation is generic, and can be used with any appropriate parameter inference problem.

The smfsbSBML package

smfsbSBML is another R package associated with the third edition of the book. This package is not on CRAN due to its dependency on a package not on CRAN, and hence is slightly less straightforward to install. Follow the available installation instructions to install the package. Once installed, you should be able to load the package with

library(smfsbSBML)

This package provides a function for reading in SBML files and parsing them into the simulatable stochastic Petri net (SPN) objects used by the main smfsb R package. Examples of suitable SBML models are included in the main smfsb GitHub repo. An appropriate SBML model can be read and parsed with a command like:

model = sbml2spn("mySbmlModel.xml")

The resulting value, model is an SPN object which can be passed in to simulation functions such as StepGillespie for constructing stochastic simulation algorithms.

Other software

In addition to the above R packages, I also have some Python scripts for converting between SBML and the SBML-shorthand notation I use in the book. See the SBML-shorthand page for further details.

Although R is a convenient language for teaching and learning about stochastic simulation, it isn’t ideal for serious research-level scientific computing or computational statistics. So for the third edition of the book I have also developed scala-smfsb, a library written in the Scala programming language, which re-implements all of the models and algorithms from the third edition of the book in Scala, a fast, efficient, strongly-typed, compiled, functional programming language. I’ll give an introduction to this library in a subsequent post, but in the meantime, it is already well documented, so see the scala-smfsb repo for further details, including information on installation, getting started, a tutorial, examples, API docs, etc.

Source

This blog post started out as an RMarkdown document, the source of which can be found here.

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Summary stats for ABC

Introduction

In the previous post I gave a very brief introduction to ABC, including a simple example for inferring the parameters of a Markov process given some time series observations. Towards the end of the post I observed that there were (at least!) two potential problems with scaling up the simple approach described, one relating to the dimension of the data and the other relating to the dimension of the parameter space. Before moving on to the (to me, more interesting) problem of the dimension of the parameter space, I will briefly discuss the data dimension problem in this post, and provide a couple of references for further reading.

Summary stats

Recall that the simple rejection sampling approach to ABC involves first sampling a candidate parameter \theta^\star from the prior and then sampling a corresponding data set x^\star from the model. This simulated data set is compared with the true data x using some (pseudo-)norm, \Vert\cdot\Vert, and accepting \theta^\star if the simulated data set is sufficiently close to the true data, \Vert x^\star - x\Vert <\epsilon. It should be clear that if we are using a proper norm then as \epsilon tends to zero the distribution of the accepted values tends to the desired posterior distribution of the parameters given the data.

However, smaller choices of \epsilon will lead to higher rejection rates. This will be a particular problem in the context of high-dimensional x, where it is often unrealistic to expect a close match between all components of x and the simulated data x^\star, even for a good choice of \theta^\star. In this case, it makes more sense to look for good agreement between particular aspects of x, such as the mean, or variance, or auto-correlation, depending on the exact problem and context. If we can find a finite set of sufficient statistics, s(x) for \theta, then it should be clear that replacing the acceptance criterion with \Vert s(x^\star) - s(x)\Vert <\epsilon will also lead to a scheme tending to the true posterior as \epsilon tends to zero (assuming a proper norm on the space of sufficient statistics), and will typically be better than the naive method, since the sufficient statistics will be of lower dimension and less “noisy” that the raw data, leading to higher acceptance rates with no loss of information.

Unfortunately for most problems of practical interest it is not possible to find low-dimensional sufficient statistics, and so people in practice use domain knowledge and heuristics to come up with a set of summary statistics, s(x) which they hope will closely approximate sufficient statistics. There is still a question as to how these statistics should be weighted or transformed to give a particular norm. This can be done using theory or heuristics, and some relevant references for this problem are given at the end of the post.

Implementation in R

Let’s now look at the problem from the previous post. Here, instead of directly computing the Euclidean distance between the real and simulated data, we will look at the Euclidean distance between some (normalised) summary statistics. First we will load some packages and set some parameters.

require(smfsb)
require(parallel)
options(mc.cores=4)
data(LVdata)
 
N=1e7
bs=1e5
batches=N/bs
message(paste("N =",N," | bs =",bs," | batches =",batches))

Next we will define some summary stats for a univariate time series – the mean, the (log) variance, and the first two auto-correlations.

ssinit <- function(vec)
{
  ac23=as.vector(acf(vec,lag.max=2,plot=FALSE)$acf)[2:3]
  c(mean(vec),log(var(vec)+1),ac23)
}

Once we have this, we can define some stats for a bivariate time series by combining the stats for the two component series, along with the cross-correlation between them.

ssi <- function(ts)
{
  c(ssinit(ts[,1]),ssinit(ts[,2]),cor(ts[,1],ts[,2]))
}

This gives a set of summary stats, but these individual statistics are potentially on very different scales. They can be transformed and re-weighted in a variety of ways, usually on the basis of a pilot run which gives some information about the distribution of the summary stats. Here we will do the simplest possible thing, which is to normalise the variance of the stats on the basis of a pilot run. This is not at all optimal – see the references at the end of the post for a description of better methods.

message("Batch 0: Pilot run batch")
prior=cbind(th1=exp(runif(bs,-6,2)),th2=exp(runif(bs,-6,2)),th3=exp(runif(bs,-6,2)))
rows=lapply(1:bs,function(i){prior[i,]})
samples=mclapply(rows,function(th){simTs(c(50,100),0,30,2,stepLVc,th)})
sumstats=mclapply(samples,ssi)
sds=apply(sapply(sumstats,c),1,sd)
print(sds)

# now define a standardised distance
ss<-function(ts)
{
  ssi(ts)/sds
}

ss0=ss(LVperfect)

distance <- function(ts)
{
  diff=ss(ts)-ss0
  sum(diff*diff)
}

Now we have a normalised distance function defined, we can proceed exactly as before to obtain an ABC posterior via rejection sampling.

post=NULL
for (i in 1:batches) {
  message(paste("batch",i,"of",batches))
  prior=cbind(th1=exp(runif(bs,-6,2)),th2=exp(runif(bs,-6,2)),th3=exp(runif(bs,-6,2)))
  rows=lapply(1:bs,function(i){prior[i,]})
  samples=mclapply(rows,function(th){simTs(c(50,100),0,30,2,stepLVc,th)})
  dist=mclapply(samples,distance)
  dist=sapply(dist,c)
  cutoff=quantile(dist,1000/N,na.rm=TRUE)
  post=rbind(post,prior[dist<cutoff,])
}
message(paste("Finished. Kept",dim(post)[1],"simulations"))

Having obtained the posterior, we can use the following code to plot the results.

th=c(th1 = 1, th2 = 0.005, th3 = 0.6)
op=par(mfrow=c(2,3))
for (i in 1:3) {
  hist(post[,i],30,col=5,main=paste("Posterior for theta[",i,"]",sep=""))
  abline(v=th[i],lwd=2,col=2)
}
for (i in 1:3) {
  hist(log(post[,i]),30,col=5,main=paste("Posterior for log(theta[",i,"])",sep=""))
  abline(v=log(th[i]),lwd=2,col=2)
}
par(op)

This gives the plot shown below. From this we can see that the ABC posterior obtained here is very similar to that obtained in the previous post using the full data. Here the dimension reduction is not that great – reducing from 32 data points to 9 summary statistics – and so the improvement in performance is not that noticable. But in higher dimensional problems reducing the dimension of the data is practically essential.

ABC Posterior

Summary and References

As before, I recommend the wikipedia article on approximate Bayesian computation for further information and a comprehensive set of references for further reading. Here I just want to highlight two references particularly relevant to the issue of summary statistics. It is quite difficult to give much practical advice on how to construct good summary statistics, but how to transform a set of summary stats in a “good” way is a problem that is reasonably well understood. In this post I did something rather naive (normalising the variance), but the following two papers describe much better approaches.

I still haven’t addressed the issue of a high-dimensional parameter space – that will be the topic of a subsequent post.

The complete R script

require(smfsb)
require(parallel)
options(mc.cores=4)
data(LVdata)
 
N=1e6
bs=1e5
batches=N/bs
message(paste("N =",N," | bs =",bs," | batches =",batches))
 
ssinit <- function(vec)
{
  ac23=as.vector(acf(vec,lag.max=2,plot=FALSE)$acf)[2:3]
  c(mean(vec),log(var(vec)+1),ac23)
}

ssi <- function(ts)
{
  c(ssinit(ts[,1]),ssinit(ts[,2]),cor(ts[,1],ts[,2]))
}

message("Batch 0: Pilot run batch")
prior=cbind(th1=exp(runif(bs,-6,2)),th2=exp(runif(bs,-6,2)),th3=exp(runif(bs,-6,2)))
rows=lapply(1:bs,function(i){prior[i,]})
samples=mclapply(rows,function(th){simTs(c(50,100),0,30,2,stepLVc,th)})
sumstats=mclapply(samples,ssi)
sds=apply(sapply(sumstats,c),1,sd)
print(sds)

# now define a standardised distance
ss<-function(ts)
{
  ssi(ts)/sds
}

ss0=ss(LVperfect)

distance <- function(ts)
{
  diff=ss(ts)-ss0
  sum(diff*diff)
}

post=NULL
for (i in 1:batches) {
  message(paste("batch",i,"of",batches))
  prior=cbind(th1=exp(runif(bs,-6,2)),th2=exp(runif(bs,-6,2)),th3=exp(runif(bs,-6,2)))
  rows=lapply(1:bs,function(i){prior[i,]})
  samples=mclapply(rows,function(th){simTs(c(50,100),0,30,2,stepLVc,th)})
  dist=mclapply(samples,distance)
  dist=sapply(dist,c)
  cutoff=quantile(dist,1000/N,na.rm=TRUE)
  post=rbind(post,prior[dist<cutoff,])
}
message(paste("Finished. Kept",dim(post)[1],"simulations"))

# plot the results
th=c(th1 = 1, th2 = 0.005, th3 = 0.6)
op=par(mfrow=c(2,3))
for (i in 1:3) {
  hist(post[,i],30,col=5,main=paste("Posterior for theta[",i,"]",sep=""))
  abline(v=th[i],lwd=2,col=2)
}
for (i in 1:3) {
  hist(log(post[,i]),30,col=5,main=paste("Posterior for log(theta[",i,"])",sep=""))
  abline(v=log(th[i]),lwd=2,col=2)
}
par(op)