"Error: Attempt to redefine node" in Mixture that changes size every iteration - bayesian
My data has three columns Time, Interval, Count. I have a mixture of Poissons that goes like this
mod_string = " model{
for(i in 2:length(Count)){
Count[i] ~ dpois(lambda.hacked[i]*z[i]+0.0001)
z[i] ~dbern(p)
lambda.hacked[i] <- mu[ clust[i] ]
Prob <- p^-(1:i) * (1-p) / p
mu <- (Time[1:i] - Interval[1:i])*lambda
clust[i] ~ dcat( Prob)
}
## Priors
lambda ~ dgamma(0.01,0.02)
p ~ dbeta(1,1)
}"
mu changes size at every iteration. As i grows, the number of clusters also grows.
How can I adapt this?
Related
How can I use fmincon() for different input parameters without using for loop?
I want to run the optimization function fmincon() over thousands of different input parameters. Briefly, the aim of the optimization is to find the optimal consumption and investment strategy that give the highest utility for a given wealth. The basic set up and functions are given as follows:
library(pracma)
library(NlcOptim)
# individual preference parameters
gamma <- 5
beta <- 0.02
Y <- 1
# financial market parameters
r <- 0.02
mu <- 0.06
sigma <- 0.2
lambda <- (mu-r)/sigma
# Merton fraction
w_star <- lambda / (gamma*sigma)
# fix random seed
set.seed(85)
scenarios <- 1000
Z_omega <- array(rnorm(scenarios,0,1), dim=c(scenarios,1)) # Brownian motion vector for E[J(W)]
# J multiple
multiple <- 1000000000
fineness <- 0.01
# define utility function
u <- function(C) {
C^(1-gamma)/(1-gamma)
}
# wealth scenario at t+1 for a given W_t
W.next <- function(W,C,fstar) {
W.tplus1 <- exp(r + fstar*sigma*lambda - 0.5*fstar^2*sigma^2 + fstar*sigma*Z_omega) * (W + Y - C)
return(W.tplus1)
}
J.simulate <- function(W.tplus1) {
floor.number <- floor((round_any(W.tplus1, fineness, f=floor) * 1/fineness)) + 1
ceiling.number <- ceiling((round_any(W.tplus1, fineness, f=ceiling) * 1/fineness)) + 1
x1 <- G_T[floor.number]
x2 <- G_T[ceiling.number]
y1 <- J_WT[floor.number]
y2 <- J_WT[ceiling.number]
# linear interpolation for J
J.tplus1.simulate <- y1 + ((W.tplus1-x1)/(x2-x1) * (y2-y1))
return(J.tplus1.simulate)
}
# define h(C,f|W)
h_t <- function(Cfstar) {
C <- Cfstar[1]
fstar <- Cfstar[2]
# wealth scenario at t+1 for a given W_t
W.tplus1 <- W.next(W,C,fstar)
# compute indirect utility for simulated W_t+1 using already compute J_WT
J.tplus1.simulate <- J.simulate(W.tplus1) # ignore wealth less than 0.001 (it can never be optimal)
# expectation of all J(W_t+1)
J_t_plus_1 <- mean(J.tplus1.simulate, na.rm=TRUE) # ignore NAs
# function h_t
indirect_utility <- log(-(u(C) + exp(-beta) * J_t_plus_1)*multiple)
return(indirect_utility)
}
For the sake of simplicity, I generated 10 wealth levels, W, to be optimized:
# wealth grid at T
G_T <- c(0.001, seq(0.01, 3, by=0.01))
J_1T <- -291331.95
J_WT <- G_T^(1-gamma) * J_1T
# wealth to be optimized
W_optim <- seq(0.01, 0.1, by=0.01)
What I did using the for loop is as follows:
# number of loop
wealth.loop <- length(W_optim)
# result vectors
C_star <- numeric(wealth.loop)
f_star <- numeric(wealth.loop)
J <- numeric(wealth.loop)
# lowerbound is fixed
lowerbound <- c(0.01,0.0001)
# optimize!
for (g in 1:wealth.loop) {
W <- W_optim[g]
x0 <- c((W+Y)*0.05,w_star) # initial input vector
upperbound <- c(W+Y-0.01,1) # upperbound depending on W
optimization <- fmincon(x0=x0, fn=h_t, lb=lowerbound, ub=upperbound, tol=1e-10)
C_star[g] <- optimization$par[1]
f_star[g] <- optimization$par[2]
J[g] <- optimization$value
print(c(g,optimization$par[1],optimization$par[2]))
}
This works well, but it takes hours to optimize over more than hundred of thousands set of different parameters. Hence, I was looking for some smarter ways of doing this, like using apply-related functions. For instance, I tried:
W <- W_optim
# input matrix
x0 <- matrix(0, nrow=length(W), ncol=2)
x0[,1] <- (W+Y)*0.05
x0[,2] <- w_star
# lowerbound the same
lowerbound <- c(0.01,0.0001)
# upperbound matrix
upperbound <- matrix(0, nrow=length(W), ncol=2)
upperbound[,1] <- W+Y-0.01
upperbound[,2] <- 1
# optimize using mapply
mapply(fmincon, x0=x0, fn=h_t, lb=lowerbound, up=upperbound)
But obviously it doesn't work. I'm not sure whether the problem is using matrix as input parameters, not vector, or I'm just using a wrong function. Is there any way to solve this problem with an efficient & smart coding?
I tried to optimize over the different parameters at once using mapply, but apparently it didn't work. Maybe I should have used another apply-related function or I should make a different structure for the input matrix?
rjags error Invalid vector argument to ilogit
I'd like to compare a betareg regression vs. the same regression using rjags
library(betareg)
d = data.frame(p= sample(c(.1,.2,.3,.4),100, replace= TRUE),
id = seq(1,100,1))
# I am looking to reproduce this regression with jags
b=betareg(p ~ id, data= d,
link = c("logit"), link.phi = NULL, type = c("ML"))
summary(b)
Below I am trying to do the same regression with rjags
#install.packages("rjags")
library(rjags)
jags_str = "
model {
#model
y ~ dbeta(alpha, beta)
alpha <- mu * phi
beta <- (1-mu) * phi
logit(mu) <- a + b*id
#priors
a ~ dnorm(0, .5)
b ~ dnorm(0, .5)
t0 ~ dnorm(0, .5)
phi <- exp(t0)
}"
id = d$id
y = d$p
model <- jags.model(textConnection(jags_str),
data = list(y=y,id=id)
)
update(model, 10000, progress.bar="none"); # Burnin for 10000 samples
samp <- coda.samples(model,
variable.names=c("mu"),
n.iter=20000, progress.bar="none")
summary(samp)
plot(samp)
I get an error on this line
model <- jags.model(textConnection(jags_str),
data = list(y=y,id=id)
)
Error in jags.model(textConnection(jags_str), data = list(y = y, id = id)) :
RUNTIME ERROR:
Invalid vector argument to ilogit
Can you advise
(1) how to fix the error
(2) how to set priors for the beta regression
Thank you.
This error occurs because you have supplied the id vector to the scalar function logit. In Jags inverse link functions cannot be vectorized. To address this, you need to use a for loop to go through each element of id. To do this I would probably add an additional element to your data list that denotes how long id is.
d = data.frame(p= sample(c(.1,.2,.3,.4),100, replace= TRUE),
id = seq(1,100,1), len_id = length(seq(1,100,1)))
From there you just need to make a small edit to your jags code.
for(i in 1:(len_id)){
y[i] ~ dbeta(alpha[i], beta[i])
alpha[i] <- mu[i] * phi
beta[i] <- (1-mu[i]) * phi
logit(mu[i]) <- a + b*id[i]
}
However, if you track mu it is going to be a matrix that is 20000 (# of iterations) by 100 (length of id). You are likely more interested in the actual parameters (a, b, and phi).
I have a code in OpenBUGS but the error is "variable CR is not defined"
model
{
for( i in 1 : N ) {
dgf[i] ~ dbin(p[i],n[i])
logit(p[i]) <- a[subject[i]] + beta[1] * CR[i]
}
for (j in 1:94)
{
a[j]~dnorm(beta0,prec.tau)
}
beta[1] ~ dnorm(0.0,.000001)
beta0 ~ dnorm(0.0,.000001)
prec.tau ~ dgamma(0.001,.001)
tau<-sqrt(1/prec.tau)
}
list(
n=c(1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1,1),
dgf=c(0,0,0,0,0,0,0,0,0,0,1,1,0,0,0,0,0,0,0,1,1,1,0,0,1,1,0,0,0,1,1,1,0,0,0,0,0,0,1,0,0,0,0,0,0,0,0,0,0,0,1,0,1,0,0,0,0,0,0,0,1,0,0,0,0,0,0,0,0,0,0,1,1,0,0,1,1,0,0,1,0,0,0,0,0,1,0,0,0,0,0,0,0,0,0,0,0,0,0,1,1,0,0,0,0,0,1,0,0,0,0,0,1,0,0,0,0,0,1,0,0,0,0,0,0,0,0,0,1,0,0,0,0,0,0,0,0,1,0,0,0,0,0,0,0,0,0,0,0,0,1,1,0,0,1,1,0,0,0,0,1,0,1,0,0,0,0,0,0,0,0,0,0,0,0,0,0,0,0,0,1,0,0,0,0,0,0,0),
subject=c(1,1,2,2,3,3,4,4,5,5,6,6,7,7,8,8,9,9,10,10,11,11,12,12,13,13,14,14,15,15,16,16,17,17,18,18,19,19,20,20,21,21,22,22,23,23,24,24,25,25,26,26,27,27,28,28,29,29,30,30,31,31,32,32,33,33,34,34,35,35,36,36,37,37,38,38,39,39,40,40,41,41,42,42,43,43,44,44,45,45,46,46,47,47,48,48,49,49,50,50,51,51,52,52,53,53,54,54,55,55,56,56,57,57,58,58,59,59,60,60,61,61,62,62,63,63,64,64,65,65,66,66,67,67,68,68,69,69,70,70,71,71,72,72,73,73,74,74,75,75,76,76,77,77,78,78,79,79,80,80,81,81,82,82,83,83,84,84,85,85,86,86,87,87,88,88,89,89,90,90,91,91,92,92,93,93,94,94),
CR=c(NA,NA,1.41,0.85,1.13,0.65,NA,NA,2.13,1.61,7.31,3.8,1.65,2.32,1.13,2.3,0.99,1.5,1.32,3.95,7.2,2.97,0.83,1.55,NA,6.5,0.89,1.2,1.52,7,8.68,7.41,NA,0.86,NA,1.92,NA,1.31,7.8,1.78,NA,1.67,NA,NA,NA,NA,NA,2.25,0.98,0.82,3.94,1.14,12,2.58,2.42,2.59,NA,NA,NA,NA,6.6,3.22,NA,2.02,2.43,1.96,0.82,1.64,1.81,1.53,1.01,5.21,8.33,1.14,1.49,6,5.6,2,3.33,4.08,NA,NA,1.14,1.25,0.85,5.42,0.85,0.65,1.02,1.33,1.1,1.12,NA,NA,1.53,1.76,2,0.85,2.9,5,4.09,2.68,0.98,1.48,0.66,0.57,5.72,2.34,0.93,2.39,1.39,1.44,4.77,2.39,1.79,1.2,0.81,1.25,4.69,1.22,1.92,1.48,2.46,NA,NA,2.53,1.12,1.74,3.45,1.22,1.27,2.61,1.75,0.82,NA,1.4,NA,5.1,1.24,1.5,1.94,1.24,1.04,1.24,NA,2.39,NA,2.07,2.19,1.6,6,6.38,1.17,1.2,5.62,6.39,1.82,1.31,NA,1.18,3.71,2.03,5.4,2.17,NA,1.94,1.57,1.44,1.35,1.63,1.24,1.54,1.5,NA,NA,NA,NA,1.44,NA,2.19,7.98,2.15,1.71,1.45,NA,0.98,2.37,1.58), N = 188)
I know that the error is because of "NA" in variable "CR" but i don't know how to solve it. i'll appreciate any help.
You have 2 options:
1) Remove the missing CR and the corresponding elements of n, dgf, and subject (and reduce N accordingly).
2) Define a stochastic relation for CR within your model, so that the model estimates the missing CR and uses these estimates in the logistic regression. Something like:
for(i in 1:N){
CR[i] ~ dnorm(CR_mu, CR_tau)
}
CR_mu ~ dnorm(0, 10^-6)
CR_tau ~ drama(0.001, 0.001)
CR_mu and CR_tau are probably not of interest but can be monitored if you want.
Note that both approaches assume that CR are missing at random (and not e.g. censored) - if CR are missing not at random this will give you biased results.
Matt
Spline in JAGS mixing badly
I have a model that calculates a spline for Mark-recapture data with survival data. The model is working fine, but the parameters that calculates the spline are mixing super badly.
mean 2.5% 97.5% That n.eff
...
m[1] 1.667899656 -0.555606 4.18479 2.8829 4
m[2] 1.293023680 -0.951046 3.90294 2.8476 4
m[3] 1.717855378 -0.484097 4.23105 2.8690 4
m[4] 1.723899423 -0.474260 4.23869 2.8686 4
m[5] 1.747050770 -0.456455 4.26314 2.8578 4
...
Basically, I'm calculating a recapture rate p composed of a species specific effect p.sp and the sampling effort p.effort. I also calculate a fitness component phi with a species specific term phi.sp, the effect of year phi.year, a climate factor phi.sum.preci and the spline m.
run.model <- function(d, ## incoming data (packaged up in src/analyses.R)
ni=1100, ## number of iterations to run ## number of draws per chain
nt=10, ## thinning rate ##to save space on computer disk space see p.61 Kéry
nb=100, ## burn in ## should be large enough to discard initial part of Markov chains that have not yet converged
nc=3, ## number of chains to run ## multiple chain to check the convergence
n.cluster = 3) {
model.jags <- function() {
## Priors ------------------------------------------------------------------
## Random effect species-specific intercept (survival)
mu.phi.sp ~ dnorm(0,0.01)
sigma.phi.sp ~ dunif(0,10)
tau.phi.sp <- 1/(sigma.phi.sp)^2
## Random effect for recapture rate
mu.p.sp ~ dnorm(0,0.01)
## Random effect of year and fixed effect of precipitation & abundance
sigma.phi.year ~ dunif(0,10)
tau.phi.year <- 1/(sigma.phi.year)^2
## fixed effect of effort
p.effort ~ dnorm(0, 0.01) ## fixed effect
## Fixed precipitation per year
phi.sum.preci ~ dnorm(0, 0.01) ## fixed effect
# Prior spline ------------------------------------------------------------
###BEGIN SPLINE###
# prior distribution for the fixed effects parameters
for (l in 1:3) {
beta[l] ~ dnorm(0,0.1)
}
prior.scaleeps <- 1
xi ~ dnorm(0, tau.xi)
tau.xi <- pow(prior.scaleeps, -2)
for (k in 1:nknotsb) {
b[k] <- yi*etab[k]
etab[k] ~ dnorm(0, tau.etab) # hierarchical model for theta
} # closing k
prior.scaleb <- 1
yi ~ dnorm (0, tau.yi)
tau.yi <- pow(prior.scaleb, -2)
tau.etab ~ dgamma(.5, .5) # chi^2 with 1 d.f.
sigmab <- abs(xi)/sqrt(tau.etab) # cauchy = normal/sqrt(chi^2)
###END SPLINE###
for(sp in 1:nsp) {
## Random species-specific intercept
phi.sp[sp] ~ dnorm(mu.phi.sp, tau.phi.sp)
## Random recapture rate
p.sp[sp] <- mu.p.sp # Changed from a comment from Luke Jan. 9 2017
}
for (yr in 1:nyear) {
## random year
phi.year[yr] ~ dnorm(0, tau.phi.year)
}
## Likelihood!
for(sp in 1:nsp) { ## per species
## Rates -------------------------------------------------------------------
## recapture rate
for (yr in 1:nyear) {
logit(p[sp,yr]) <- # added logit here
p.sp[sp] +
p.effort*effort[yr]
} ## closing for (year in 1:nyear)
} ## closing for (sp in 1:nsp)
## Each ID ----------------------------------------------------------------
## Likelihood!
for(ind in 1:nind) { ## nind = nrow(d$X)
### BEGIN SPLINE ###
## mean function model
m[ind] <-mfe[ind] + mre1[ind] + mre2[ind]
# fixed effect part
mfe[ind] <- beta[1] * Xfix[ind,1] +beta[2] * Xfix[ind,2] + beta[3] * Xfix[ind,3]
mre1[ind] <- b[1]*Z[ind,1] + b[2]*Z[ind,2] + b[3]*Z[ind,3] + b[4]*Z[ind,4] + b[5]*Z[ind,5] + b[6]*Z[ind,6] + b[7]*Z[ind,7] + b[8]*Z[ind,8] + b[9]*Z[ind,9] + b[10]*Z[ind,10]
mre2[ind] <- b[11]*Z[ind,11] + b[12]*Z[ind,12] + b[13]*Z[ind,13] + b[14]*Z[ind,14] + b[15]*Z[ind,15]
###END SPLINE###
}
## for each individual
for(ind in 1:nind) { ## nind = nrow(d$X)
for(yr in 1:nyear) {
logit(phi[ind,yr]) <-
phi.sp[species[ind]] + ## effect of species
phi.year[yr] + ## effect of year
# Effect of the traits on survival values
m[ind]+ # spline
phi.sum.preci*sum.rainfall[yr] # effect of precipitation per sampling event
} ## (yr in 1:nyear)
## First occasion
for(yr in 1:first[ind]) {
z[ind,yr] ~ dbern(1)
} ## (yr in 1:first[ind])
## Subsequent occasions
for(yr in (first[ind]+1):nyear) { # (so, here, we're just indexing from year "first+1" onwards).
mu.z[ind,yr] <- phi[ind,yr-1]*z[ind,yr-1]
z[ind,yr] ~ dbern(mu.z[ind,yr])
## Observation process
sight.p[ind,yr] <- z[ind,yr]*p[species[ind],yr] ## sightp probability of something to be seen
X[ind,yr] ~ dbern(sight.p[ind,yr]) ## X matrix : ind by years
} ## yr
} ## closing for(ind in 1:nind)
} ## closing model.jags function
## Calling Jags ------------------------------------------------------------
jags.parallel(data = d$data,
inits = d$inits,
parameters.to.save = d$params,
model.file = model.jags,
n.chains = nc, n.thin = nt, n.iter = ni, n.burnin = nb,
working.directory = NULL,
n.cluster = n.cluster)
} ## closing the run.model function
# Monitored parameters ----------------------------------------------------
get.params <- function()
c('phi.sp','mu.phi.sp','sigma.phi.sp','mu.p.sp','sigma.p.sp','phi.year','phi','p', 'phi.sum.preci','p.sp','p.effort','z',
# Spline parameters
"m","sigmab","b","beta")
Why the jags result and depmixS4 are sometimes different?
I have a data set like the following simulated data:
Pi = matrix(c(0.9,0.1,0.3,0.7),2,2,byrow=TRUE)
delta = c(.5,.5)
z = sample(c(1,2),1,prob=delta)
T = 365
for( t in 2:T){
z[t] = sample(x=c(1,2),1,prob=Pi[z[t-1],])
}
x <- sample(x=seq(-1, 1.5, length.out=T),T,replace=TRUE)
alpha = c(-1, -3.2)
Beta = c(-4,3)
y<-NA
for(i in 1:T){
y[i] = rbinom(1,size=10,prob=1/(1+exp(-Beta[z[i]]*x[i]-alpha[z[i]])))
}
SimulatedBinomData <- data.frame('y' = y, 'x' = x , size=rep(10,T), 'z' = z)
yy<-NA
xx<-NA
for(i in 1:dim(SimulatedBinomData)[1]){
yy<-c(yy,c(rep(1,SimulatedBinomData$y[i]),rep(0,(SimulatedBinomData$size[i]-SimulatedBinomData$y[i]))))
xx<-c(xx,rep(SimulatedBinomData$x[i],SimulatedBinomData$size[i]))
}
yy<-yy[-1]
xx<-xx[-1]
SimulatedBernolliData<-data.frame(y=yy,x=xx, tt=rep(c(1:T),rep(10,T)))
This is a HMM problem with two states meaning that the Hidden Markov chain z_t belongs to {1,2}. To estimate alpha and Beta in two different states I can use the package 'depmixS4' and find the Maximum Likelihood estimates or I can use MCMC in 'rjags' package.
I expect that these two estimations be almost the same while when I run the following program for different simulated data, in several times, the answers are not the same and very different!!
library("rjags")
library("depmixS4")
mod <- depmix(cbind(y,(size-y))~x, data=SimulatedBinomData, nstates=2, family=binomial(logit))
fm <- fit(mod)
getpars(fm)
n<-length(SimulatedBernolliData$y)
T<-max(SimulatedBernolliData$tt)
cat("model {
# Transition Probability
Ptrans[1,1:2] ~ ddirch(a)
Ptrans[2,1:2] ~ ddirch(a)
# States
Pinit[1] <- 0.5 #failor
Pinit[2] <- 0.5 #success
state[1] ~ dbern(Pinit[2])
for (t in 2:T) {
state[t] ~ dbern(Ptrans[(state[t-1]+1),2])
}
# Parameters
alpha[1] ~ dunif(-1.e10, 1.e10)
alpha[2] ~ dunif(-1.e10, 1.e10)
Beta[1] ~ dunif(-1.e10, 1.e10)
Beta[2] ~ dunif(-1.e10, 1.e10)
# Observations
for (i in 1:n){
z[i] <- state[tt[i]]
y[i] ~ dbern(1/(1+exp(-(alpha[(z[i]+1)]+Beta[(z[i]+1)]*x[i]))))
}
}",
file="LeftBehindHiddenMarkov.bug")
jags <- jags.model('LeftBehindHiddenMarkov.bug', data = list('x' = SimulatedBernolliData$x, 'y' = SimulatedBernolliData$y, 'tt' = SimulatedBernolliData$tt, T=T, n = n, a = c(1,1) ))
res <- coda.samples(jags,c('alpha', 'Beta', 'Ptrans','state'),1000)
res.median = apply(res[[1]],2,median)
res.median[1:8]
res.mean = apply(res[[1]],2,mean)
res.mean[1:8]
res.sd = apply(res[[1]],2,sd)
res.sd[1:8]
res.mode = apply(res[[1]],2,function(x){as.numeric(names(table(x))
[which.max(table(x))]) })
res.mode[1:8]
You are having a problem of label switching in your JAGS code, that is, states z[i]=1 is not bounded to the lower posterior value for Beta and z[i]=2 to the higher Beta. Therefore, for each iteration of the MCMC they can switch. There are several ways to solve this problem. One of them is the partial reordering, that is, for every MCMC iteration, draw two independent values for Beta and order them so that Beta[1] < Beta[2]. You can do that by substituting Beta[1] ~ dunif(-1.e10, 1.e10) Beta[2] ~ dunif(-1.e10, 1.e10) for Beta[1:2] <- sort(Betaaux) Betaaux[1] ~ dunif(-1.e10, 1.e10) Betaaux[2] ~ dunif(-1.e10, 1.e10) Of course, the ordering could also be done on the alpha parameters instead. The election of which parameter to use for the partial reordering depends on the problem.