In this vignette, we explain how one can compute marginal likelihoods, Bayes factors, and posterior model probabilities using a simple hierarchical normal model implemented in `JAGS`

. This vignette uses the same models and data as the `Stan`

vignette.

The model that we will use assumes that each of the \(n\) observations \(y_i\) (where \(i\) indexes the observation, \(i = 1,2,...,n\)) is normally distributed with corresponding mean \(\theta_i\) and a common known variance \(\sigma^2\): \(y_i \sim \mathcal{N}(\theta_i, \sigma^2)\). Each \(\theta_i\) is drawn from a normal group-level distribution with mean \(\mu\) and variance \(\tau^2\): \(\theta_i \sim \mathcal{N}(\mu, \tau^2)\). For the group-level mean \(\mu\), we use a normal prior distribution of the form \(\mathcal{N}(\mu_0, \tau^2_0)\). For the group-level variance \(\tau^2\), we use an inverse-gamma prior of the form \(\text{Inv-Gamma}(\alpha, \beta)\). We will use `JAGS`

to fit the model which parametrizes the normal distribution in terms of the precision (i.e., one over the variance). Consequently, we implement this inverse-gamma prior on \(\tau^2\) by placing a gamma prior of the form \(\text{Gamma}(\alpha, \beta)\) on the precision; we call this precision parameter `invTau2`

in the code.

In this example, we are interested in comparing the null model \(\mathcal{H}_0\), which posits that the group-level mean \(\mu = 0\), to the alternative model \(\mathcal{H}_1\), which allows \(\mu\) to be different from zero. First, we generate some data from the null model:

```
library(bridgesampling)
### generate data ###
set.seed(12345)
<- 0
mu <- 0.5
tau2 <- 1
sigma2
<- 20
n <- rnorm(n, mu, sqrt(tau2))
theta <- rnorm(n, theta, sqrt(sigma2)) y
```

Next, we specify the prior parameters \(\mu_0\), \(\tau^2_0\), \(\alpha\), and \(\beta\):

```
### set prior parameters ###
<- 0
mu0 <- 1
tau20 <- 1
alpha <- 1 beta
```

Now we can fit the null and the alternative model in `JAGS`

(note that it is necessary to install `JAGS`

for this). One usually requires a larger number of posterior sample for estimating the marginal likelihood than for simply estimating the model parameters. This is the reason for using a comparatively large number of samples (i.e., 50,000 post burn-in samples per chain) for this comparatively simple model.

```
library(R2jags)
### functions to get posterior samples ###
# H0: mu = 0
<- function(data, niter = 52000, nburnin = 2000, nchains = 3) {
getSamplesModelH0
<- "
model model {
for (i in 1:n) {
theta[i] ~ dnorm(0, invTau2)
y[i] ~ dnorm(theta[i], 1/sigma2)
}
invTau2 ~ dgamma(alpha, beta)
tau2 <- 1/invTau2
}"
<- jags(data, parameters.to.save = c("theta", "invTau2"),
s model.file = textConnection(model),
n.chains = nchains, n.iter = niter,
n.burnin = nburnin, n.thin = 1)
return(s)
}
# H1: mu != 0
<- function(data, niter = 52000, nburnin = 2000,
getSamplesModelH1 nchains = 3) {
<- "
model model {
for (i in 1:n) {
theta[i] ~ dnorm(mu, invTau2)
y[i] ~ dnorm(theta[i], 1/sigma2)
}
mu ~ dnorm(mu0, 1/tau20)
invTau2 ~ dgamma(alpha, beta)
tau2 <- 1/invTau2
}"
<- jags(data, parameters.to.save = c("theta", "mu", "invTau2"),
s model.file = textConnection(model),
n.chains = nchains, n.iter = niter,
n.burnin = nburnin, n.thin = 1)
return(s)
}
### get posterior samples ###
# create data lists for JAGS
<- list(y = y, n = length(y), alpha = alpha, beta = beta, sigma2 = sigma2)
data_H0 <- list(y = y, n = length(y), mu0 = mu0, tau20 = tau20, alpha = alpha,
data_H1 beta = beta, sigma2 = sigma2)
# fit models
<- getSamplesModelH0(data_H0)
samples_H0 <- getSamplesModelH1(data_H1) samples_H1
```

The next step is to write the corresponding `log_posterior`

(i.e., unnormalized posterior) function for both models. This function takes one draw from the joint posterior and the data object as input and returns the log of the unnormalized joint posterior density. When using MCMC software such as `JAGS`

or `Stan`

, specifying this function is relatively simple. As a rule of thumb, one only needs to look for all places where a “`~`

” sign appears in the model code. The log of the densities on the right-hand side of these “`~`

” symbols needs to be evaluated for the relevant quantities and then these log densities values are summed.

For example, in the null model, there are three “`~`

” signs. Starting at the data-level, we need to evaluate the log of the normal density with mean \(\theta_i\) and variance \(\sigma^2\) for all \(y_i\) and then sum the resulting log density values. Next, we move one step up in the model and evaluate the log of the group-level density for all \(\theta_i\). Hence, we evaluate the log of the normal density for \(\theta_i\) with mean \(\mu = 0\) and variance \(\tau^2\) (remember that `JAGS`

parametrizes the normal distribution in terms of the precision `invTau2`

= \(1/\tau^2\); in contrast, `R`

parametrizes it in terms of the standard deviation) and sum the resulting log density values. The result of this summation is added to the result of the previous summation for the data-level normal distribution. Finally, we need to evaluate the log of the prior density for `invTau2`

. This means that we compute the log density of the gamma distribution with parameters \(\alpha\) and \(\beta\) for the sampled `invTau2`

value and add the resulting log density value to the result of summing the data-level and group-level log densities. The unnormalized log posterior for the alternative model can be obtained in a similar fashion. The resulting functions look as follows:

```
### functions for evaluating the unnormalized posteriors on log scale ###
<- function(samples.row, data) {
log_posterior_H0
<- 0
mu <- samples.row[[ "invTau2" ]]
invTau2 <- samples.row[ paste0("theta[", seq_along(data$y), "]") ]
theta
sum(dnorm(data$y, theta, data$sigma2, log = TRUE)) +
sum(dnorm(theta, mu, 1/sqrt(invTau2), log = TRUE)) +
dgamma(invTau2, data$alpha, data$beta, log = TRUE)
}
<- function(samples.row, data) {
log_posterior_H1
<- samples.row[[ "mu" ]]
mu <- samples.row[[ "invTau2" ]]
invTau2 <- samples.row[ paste0("theta[", seq_along(data$y), "]") ]
theta
sum(dnorm(data$y, theta, data$sigma2, log = TRUE)) +
sum(dnorm(theta, mu, 1/sqrt(invTau2), log = TRUE)) +
dnorm(mu, data$mu0, sqrt(data$tau20), log = TRUE) +
dgamma(invTau2, data$alpha, data$beta, log = TRUE)
}
```

The final step before computing the log marginal likelihoods is to specify the parameter bounds. In this example, for both models, all parameters can range from \(-\infty\) to \(\infty\) except the precision `invTau2`

which has a lower bound of zero. These boundary vectors need to be named and the names need to match the order of the parameters.

```
# specify parameter bounds H0
<- colnames(samples_H0$BUGSoutput$sims.matrix)
cn <- cn[cn != "deviance"]
cn <- rep(-Inf, length(cn))
lb_H0 <- rep(Inf, length(cn))
ub_H0 names(lb_H0) <- names(ub_H0) <- cn
"invTau2" ]] <- 0
lb_H0[[
# specify parameter bounds H1
<- colnames(samples_H1$BUGSoutput$sims.matrix)
cn <- cn[cn != "deviance"]
cn <- rep(-Inf, length(cn))
lb_H1 <- rep(Inf, length(cn))
ub_H1 names(lb_H1) <- names(ub_H1) <- cn
"invTau2" ]] <- 0 lb_H1[[
```

Note that currently, the lower and upper bound of a parameter cannot be a function of the bounds of another parameter. Furthermore, constraints that depend on multiple parameters of the model are not supported. This excludes, for example, parameters that constitute a covariance matrix or sets of parameters that need to sum to one.

Now we are ready to compute the log marginal likelihoods using the `bridge_sampler`

function. We use `silent = TRUE`

to suppress printing the number of iterations to the console:

```
# compute log marginal likelihood via bridge sampling for H0
<- bridge_sampler(samples = samples_H0, data = data_H0,
H0.bridge log_posterior = log_posterior_H0, lb = lb_H0,
ub = ub_H0, silent = TRUE)
# compute log marginal likelihood via bridge sampling for H1
<- bridge_sampler(samples = samples_H1, data = data_H1,
H1.bridge log_posterior = log_posterior_H1, lb = lb_H1,
ub = ub_H1, silent = TRUE)
```

We obtain:

`print(H0.bridge)`

```
## Bridge sampling estimate of the log marginal likelihood: -37.53235
## Estimate obtained in 4 iteration(s) via method "normal".
```

`print(H1.bridge)`

```
## Bridge sampling estimate of the log marginal likelihood: -37.79776
## Estimate obtained in 5 iteration(s) via method "normal".
```

We can use the `error_measures`

function to compute an approximate percentage error of the estimates:

```
# compute percentage errors
<- error_measures(H0.bridge)$percentage
H0.error <- error_measures(H1.bridge)$percentage H1.error
```

We obtain:

`print(H0.error)`

`## [1] "0.144%"`

`print(H1.error)`

`## [1] "0.162%"`

To compare the null model and the alternative model, we can compute the Bayes factor by using the `bf`

function. In our case, we compute \(\text{BF}_{01}\), that is, the Bayes factor which quantifies how much more likely the data are under the null versus the alternative model:

```
# compute Bayes factor
<- bf(H0.bridge, H1.bridge)
BF01 print(BF01)
```

`## Estimated Bayes factor in favor of H0.bridge over H1.bridge: 1.30396`

In this case, the Bayes factor is close to one, indicating that there is not much evidence for either model. We can also compute posterior model probabilities by using the `post_prob`

function:

```
# compute posterior model probabilities (assuming equal prior model probabilities)
<- post_prob(H0.bridge, H1.bridge)
post1 print(post1)
```

```
## H0.bridge H1.bridge
## 0.5659652 0.4340348
```

When the argument `prior_prob`

is not specified, as is the case here, the prior model probabilities of all models under consideration are set equal (i.e., in this case with two models to 0.5). However, if we had prior knowledge about how likely both models are, we could use the `prior_prob`

argument to specify different prior model probabilities:

```
# compute posterior model probabilities (using user-specified prior model probabilities)
<- post_prob(H0.bridge, H1.bridge, prior_prob = c(.6, .4))
post2 print(post2)
```

```
## H0.bridge H1.bridge
## 0.6616986 0.3383014
```