---
title: "Visual MCMC diagnostics using the bayesplot package"
author: "Jonah Gabry and Martin Modrák"
date: "`r Sys.Date()`"
output: 
  rmarkdown::html_vignette:
    toc: true
    toc_depth: 3
params:
  EVAL: !r identical(Sys.getenv("NOT_CRAN"), "true")
vignette: >
  %\VignetteIndexEntry{Visual MCMC diagnostics}
  %\VignetteEngine{knitr::rmarkdown}
  %\VignetteEncoding{UTF-8}
---

```{r, child="children/SETTINGS-knitr.txt"}
```

```{r, pkgs, include=FALSE}
library("bayesplot")
library("ggplot2")
library("rstan")
library("dplyr") #Used only for consistency checks
rstan_options(auto_write = TRUE) #Helpful throughout development
```

## Introduction

This vignette focuses on MCMC diagnostic plots, in particular on diagnosing 
divergent transitions and on the `n_eff` and `Rhat` statistics that help you
determine that the chains have mixed well. Plots of parameter estimates
from MCMC draws are covered in the separate vignette
[_Plotting MCMC draws_](http://mc-stan.org/bayesplot/articles/plotting-mcmc-draws.html), 
and graphical posterior predictive model checking is covered in the
[_Graphical posterior predictive checks_](http://mc-stan.org/bayesplot/articles/graphical-ppcs.html)
vignette.

Note that most of these plots can also be browsed interactively using the 
[shinystan](http://mc-stan.org/shinystan/) package.

### Setup

In addition to __bayesplot__ we'll load the following packages: 

* __ggplot2__, in case we want to customize the ggplot objects created by __bayesplot__
* __rstan__, for fitting the example models used throughout the vignette

```{r, eval=FALSE}
library("bayesplot")
library("ggplot2")
library("rstan")      
```

### Example model

Before we delve into the actual plotting we need to fit a model to have something to work with.
In this vignette we'll use the eight schools example, which is discussed in many 
places, including Rubin (1981), Gelman et al. (2013), and the 
[RStan Getting Started](https://github.com/stan-dev/rstan/wiki/RStan-Getting-Started#how-to-use-rstan)
wiki. This is a simple hierarchical meta-analysis model with data consisting of 
point estimates `y` and standard errors `sigma` from analyses of test prep 
programs in `J=8` schools. Ideally we would have the full data from each of 
the previous studies, but in this case we only have the these estimates.

```{r, schools_dat}
schools_dat <- list(
  J = 8, 
  y = c(28,  8, -3,  7, -1,  1, 18, 12),
  sigma = c(15, 10, 16, 11,  9, 11, 10, 18)
)
```

The model is: 
$$
\begin{align*}
y_j &\sim {\rm Normal}(\theta_j, \sigma_j), \quad j = 1,\dots,J \\
\theta_j &\sim {\rm Normal}(\mu, \tau), \quad j = 1, \dots, J \\
\mu &\sim {\rm Normal}(0, 10) \\
\tau &\sim {\rm half-Cauchy}(0, 10),
\end{align*}
$$
with the normal distribution parameterized by the mean and standard deviation, 
not the variance or precision. In Stan code:
```{stan, stancode1, output.var = "schools_mod_cp"}
// Saved in 'schools_mod_cp.stan'
data {
  int<lower=0> J;
  vector[J] y;
  vector<lower=0>[J] sigma;
}
parameters {
  real mu;
  real<lower=0> tau;
  vector[J] theta;
}
model {
  mu ~ normal(0, 10);
  tau ~ cauchy(0, 10);
  theta ~ normal(mu, tau);
  y ~ normal(theta, sigma);
}
```

This parameterization of the model is referred to as the centered 
parameterization (CP). We'll also fit the same statistical model but using the 
so-called non-centered parameterization (NCP), which replaces the vector 
$\theta$ with a vector $\eta$ of a priori _i.i.d._ standard normal parameters 
and then constructs $\theta$ deterministically from $\eta$ by scaling by $\tau$ 
and shifting by $\mu$:
$$
\begin{align*}
\theta_j &= \mu + \tau \,\eta_j, \quad j = 1,\dots,J \\
\eta_j &\sim N(0,1), \quad j = 1,\dots,J.
\end{align*}
$$
The Stan code for this model is:
```{stan, stancode2, output.var = "schools_mod_ncp"}
// Saved in 'schools_mod_ncp.stan'
data {
  int<lower=0> J;
  vector[J] y;
  vector<lower=0>[J] sigma;
}
parameters {
  real mu;
  real<lower=0> tau;
  vector[J] eta;
}
transformed parameters {
  vector[J] theta;
  theta = mu + tau * eta;
}
model {
  mu ~ normal(0, 10);
  tau ~ cauchy(0, 10);
  eta ~ normal(0, 1); // implies theta ~ normal(mu, tau)
  y ~ normal(theta, sigma);
}
```

The centered and non-centered are two parameterizations of the same statistical 
model, but they have very different practical implications for MCMC. Using the 
__bayesplot__ diagnostic plots, we'll see that, for this data, the NCP is 
required in order to properly explore the posterior distribution.

To fit both models we first translate the Stan code to C++ and compile it using 
the `stan_model` function.
```{r, compile-models, eval=FALSE}
schools_mod_cp <- stan_model("schools_mod_cp.stan")
schools_mod_ncp <- stan_model("schools_mod_ncp.stan")
```
We then fit the model by calling Stan's MCMC algorithm using the `sampling` 
function (the increased `adapt_delta` param is to make the sampler a bit more "careful" and avoid false positive divergences),
```{r, fit-models-hidden, results='hide', message=FALSE}
fit_cp <- sampling(schools_mod_cp, data = schools_dat, seed = 803214054, control = list(adapt_delta = 0.95))
fit_ncp <- sampling(schools_mod_ncp, data = schools_dat, seed = 457721433, control = list(adapt_delta = 0.95))
```
and extract a `iterations x chains x parameters` array of posterior draws with 
`as.array`,
```{r, extract-draws}
# Extract posterior draws for later use
posterior_cp <- as.array(fit_cp)
posterior_ncp <- as.array(fit_ncp)
```

You may have noticed the warnings about divergent transitions for the centered
parametrization fit. Those are serious business and in most cases indicate that
something is wrong with the model and the results should not be trusted. 
For an explanation of these warnings see 
[Divergent transitions after warmup](http://mc-stan.org/misc/warnings.html#divergent-transitions-after-warmup).
We'll have a look at diagnosing the source of the divergences first and then
dive into some diagnostics that should be checked even if there are no warnigns
from the sampler.

<br>

## Diagnostics for the No-U-Turn Sampler

The No-U-Turn Sampler (NUTS, Hoffman and Gelman, 2014) is the variant of 
Hamiltonian Monte Carlo (HMC) used by [Stan](http://mc-stan.org/) and the
various R packages that depend on Stan for fitting Bayesian models.
The **bayesplot** package has special functions for visualizing some of the 
unique diagnostics permitted by HMC, and NUTS in particular. See Betancourt 
(2017), Betancourt and Girolami (2013), and Stan Development Team (2017) for
more details on the concepts.

**Documentation:**

* `help("MCMC-nuts")` 
* [mc-stan.org/bayesplot/reference/MCMC-nuts](http://mc-stan.org/bayesplot/reference/MCMC-nuts.html)

------


The special **bayesplot** functions for NUTS diagnostics are

```{r, available_mcmc-nuts}
available_mcmc(pattern = "_nuts_")
```


Those functions require more information than simply the posterior draws, in
particular the log of the posterior density for each draw and some
NUTS-specific diagnostic values may be needed. The **bayesplot** package
provides generic functions `log_posterior` and `nuts_params` for extracting this
information from fitted model objects. Currently methods are provided for models
fit using the **rstan**, **rstanarm** and **brms** packages, although it is not
difficult to define additional methods for the objects returned by other R
packages. For the Stan models we fit above we can use the `log_posterior` and
`nuts_params` methods for stanfit objects:

```{r, extract-nuts-info}
lp_cp <- log_posterior(fit_cp)
head(lp_cp)
np_cp <- nuts_params(fit_cp)
head(np_cp)

# for the second model
lp_ncp <- log_posterior(fit_ncp)
np_ncp <- nuts_params(fit_ncp)
```

```{r echo=FALSE, warning=FALSE}
# On rare ocasions, the fits may not be illustrative. Currently the seed is
# fixed, but if something in Stan changes and the fixed seeds produce unexpected
# results (which sould be rare), we want to know.
n_divergent_cp <- np_cp %>% filter(Parameter == "divergent__" & Value == 1) %>% nrow()
n_divergent_ncp <- np_ncp %>% filter(Parameter == "divergent__" & Value == 1) %>% nrow()

if(n_divergent_cp < 10 || n_divergent_cp > 2000) {
  stop("Unexpected number of divergences in the CP model. Change seed?")
}
if(n_divergent_ncp > 0) {
  stop("Divergences in the NCP model. Fix a bug / change seed?")
}

```

In addition to the NUTS-specific plotting functions, some of the general 
MCMC plotting functions demonstrated in the 
[_Plotting MCMC draws_](http://mc-stan.org/bayesplot/articles/plotting-mcmc-draws.html)
vignette also take optional arguments that can be used to display important
HMC/NUTS diagnostic information. We'll see examples of this in the next
section on divergenent transitions.


### Divergent transitions

When running the Stan models above, there were warnings about divergent
transitions. Here we'll look at diagnosing the source of divergences through
visualizations.

#### mcmc_parcoord

The parallel coordinates plot (`mcmc_parcoord`) is probably the first plot to
have look at if you have no idea, where the divergences in your model might be
coming from. This function works in general without including information about
the divergences, but if the optional `np` argument is used to pass NUTS
parameter information, then divergences will be colored in the plot (by default
in red).

```{r, mcmc_parcoord-1}
color_scheme_set("darkgray")
mcmc_parcoord(posterior_cp, np = np_cp)
```

The `mcmc_parcoord` plot shows one line per iteration, connecting the parameter
values at this iteration. This lets you see any global patterns in the
divergences. Here, you may notice that divergences in the centered
parameterization happen exclusively when  `tau`, the hierarchical standard
deviation, goes near zero and the values of the `theta`s are essentially fixed.
This makes `tau` immediately suspect. 
See [Gabry et al. (2018)](https://arxiv.org/abs/1709.01449) for another example
of the parallel coordinates plot.


#### mcmc_pairs

The `mcmc_pairs` function can be also be used to look at multiple
parameters at once, but unlike `mcmc_parcoord` (which works well even when
including several dozen parameters) `mcmc_pairs` is more useful for up to ~8
parameters. It shows univariate histograms and bivariate scatter plots for
selected parameters and is especially useful in identifying collinearity between
variables (which manifests as narrow bivariate plots) as well as the presence of
multiplicative non-identifiabilities (banana-like shapes). 

Let's look at how `tau` interacts with other variables, using only one of the
`theta`s to keep the plot readable:

```{r, mcmc_pairs}
mcmc_pairs(posterior_cp, np = np_cp, pars = c("mu","tau","theta[1]"))
```

Note that each bivariate plot is present twice -- by default each of those
contain half of the chains, so you also get to see if the chains produced
similar results (see the documentation for the `condition` argument for 
other options). Here, the interaction of `tau` and `theta[1]` seems most
interesting, as it concentrates the divergences into a tight region.

Further examples of pairs plots and instructions for using the various optional
arguments to `mcmc_pairs` are provided via `help("mcmc_pairs")`.


#### mcmc_scatter

Using the `mcmc_scatter` function (with optional argument `np`) we can look at a
single bivariate plot to investigate it more closely. For hierarchical models, 
a good place to start is to plot a "local" parameter (`theta[j]`) 
against a "global" scale parameter on which it depends (`tau`). 

We will also use the `transformations` argument to look at the log of `tau`, as
this is what Stan is doing under the hood for parameters like `tau` that have a
lower bound of zero. That is, even though the draws for `tau` returned from
Stan are all positive, the parameter space that the Markov chains actual explore
is unconstrained. Transforming `tau` is not strictly necessary for the plot
(often the plot is still useful without it) but plotting in the unconstrained is
often even more informative.

First the plot for the centered parameterization:

```{r, mcmc_scatter-1}
# assign to an object so we can reuse later
scatter_theta_cp <- mcmc_scatter(
  posterior_cp, 
  pars = c("theta[1]", "tau"), 
  transform = list(tau = "log"), # can abbrev. 'transformations'
  np = np_cp
)
scatter_theta_cp
```

The shape of this bivariate distribution resembles a funnel (or tornado). 
This one in particular is essentially the same as an example referred to as 
Neal's funnel (details in the Stan manual) and it is a clear indication that 
the Markov chains are struggling to explore the tip of the funnel, which 
is narrower than the rest of the space. 

The main problem is that large steps are required to explore the less narrow
regions efficiently, but those steps become too large for navigating the narrow
region. The required step size is connected to the value of `tau`. When `tau` is
large it allows for large variation in `theta` (and requires large steps) while
small `tau` requires small steps in `theta`.

The non-centered parameterization avoids this by sampling the `eta` parameter
which, unlike `theta`, is _a priori independent_ of `tau`. Then `theta` is
computed deterministically from the parameters `eta`, `mu` and `tau` afterwards.
Here's the same plot as above, but with `eta[1]` from non-centered
parameterization instead of `theta[1]` from the centered parameterization:

```{r, mcmc_scatter-2}
scatter_eta_ncp <- mcmc_scatter(
  posterior_ncp, 
  pars = c("eta[1]", "tau"), 
  transform = list(tau = "log"), 
  np = np_ncp
)
scatter_eta_ncp
```

We can see that the funnel/tornado shape is replaced by a somewhat Gaussian
blob/cloud and the divergences go away. Gabry et al. (2018) has further
discussion of this example.

Ultimately we only care about `eta` insofar as it enables the Markov chains to
better explore the posterior, so let's directly examine how much more
exploration was possible after the reparameterization. For the non-centered
parameterization we can make the same scatterplot but use the values of
`theta[1] = mu + eta[1] * tau` instead of `eta[1]`. Below is a side by side
comparison with the scatterplot of `theta[1]` vs `log(tau)` from the centered
parameterization that we made above. We will also force the plots to have the
same $y$-axis limits, which will make the most important difference much more
apparent:

```{r, mcmc_scatter-3}
# A function we'll use several times to plot comparisons of the centered 
# parameterization (cp) and the non-centered parameterization (ncp). See
# help("bayesplot_grid") for details on the bayesplot_grid function used here.
compare_cp_ncp <- function(cp_plot, ncp_plot, ncol = 2, ...) {
  bayesplot_grid(
    cp_plot, ncp_plot, 
    grid_args = list(ncol = ncol),
    subtitles = c("Centered parameterization", 
                  "Non-centered parameterization"),
    ...
  )
}


scatter_theta_ncp <- mcmc_scatter(
  posterior_ncp, 
  pars = c("theta[1]", "tau"), 
  transform = list(tau = "log"), 
  np = np_ncp
)

compare_cp_ncp(scatter_theta_cp, scatter_theta_ncp, ylim = c(-9, 4))
```

Once we transform the `eta` values into `theta` values we actually see an even
more pronounced funnel/tornado shape than we have with the centered
parameterization. But this is precisely what we want! The non-centered
parameterization allowed us to obtain draws from the funnel distribution
without having to directly navigate the curvature of the funnel. With the
centered parameterization the chains never could make it into the neck of funnel
and we see a clustering of divergences and no draws in the tail of the
distribution.



#### mcmc_trace

Another useful diagnostic plot is the trace plot, which is a time series plot
of the Markov chains. That is, a trace plot shows the evolution of parameter
vector over the iterations of one or many Markov chains. The `np` argument to
the `mcmc_trace` function can be used to add a rug plot of the divergences to a
trace plot of parameter draws. Typically we can see that at least one of the
chains is getting stuck wherever there is a cluster of many red marks.

Here is the trace plot for the `tau` parameter from the centered parameterization:
```{r, mcmc_trace}
color_scheme_set("mix-brightblue-gray")
mcmc_trace(posterior_cp, pars = "tau", np = np_cp) + 
  xlab("Post-warmup iteration")
```

The first thing to note is that all chains seem to be exploring the same region
of parameter values, which is a good sign. But the plot is too crowded to help
us diagnose divergences. We may however zoom in to investigate, using the
`window` parameter:

```{r echo=FALSE}
#A check that the chosen window still relevant
n_divergent_in_window <- np_cp %>% filter(Parameter == "divergent__" & Value == 1 & Iteration >= 350 & Iteration <= 500) %>% nrow()

if(n_divergent_in_window < 6) {
  divergences <- np_cp %>% filter(Parameter == "divergent__" & Value == 1) %>% select(Iteration) %>% get("Iteration", .) %>% sort() %>% paste(collapse = ",")
  stop(paste("Too few divergences in the selected window for traceplot zoom. Change the window or the random seed.\nDivergences happened at: ", divergences))
}
```

```{r, mcmc_trace_zoom}
mcmc_trace(posterior_cp, pars = "tau", np = np_cp, window = c(400,550)) + 
  xlab("Post-warmup iteration")
```

What we see here is that chains can get stuck as `tau` approaches zero and
spend substantial time in the same region of the parameter space. This is just
another indication that there is problematic geometry at $\tau \simeq 0$ --
healthy chains jump up and down frequently.

#### mcmc_nuts_divergence

To understand how the divergences interact with the model globally, we can use
the `mcmc_nuts_divergence` function:

```{r, mcmc_nuts_divergence}
color_scheme_set("red")
mcmc_nuts_divergence(np_cp, lp_cp)
```

In the top panel we see the distribution of the log-posterior when there was no 
divergence vs the distribution when there was a divergence. Divergences often 
indicate that some part of the posterior isn't being explored and the plot 
confirms that `lp|Divergence` indeed has lighter tails than `lp|No divergence`.

The bottom panel shows the same thing but instead of the log-posterior the NUTS 
acceptance statistic is shown.

Specifying the optional `chain` argument will overlay the plot just for a
particular Markov chain on the plot for all chains combined:

```{r, mcmc_nuts_divergence-chain}
mcmc_nuts_divergence(np_cp, lp_cp, chain = 4)
```

For the non-centered parameterization we may get a few warnings about
divergences but if we do we'll have far fewer of them to worry about.

```{r, mcmc_nuts_divergence-2}
mcmc_nuts_divergence(np_ncp, lp_ncp)
```

If there are only a few divergences we can often get rid of them by increasing
the target acceptance rate (`adapt_delta`, the upper limit is 1), which has the effect of lowering the
step size used by the sampler and allowing the Markov chains to explore more
complicated curvature in the target distribution.

```{r, fit-adapt-delta, results='hide', message=FALSE}
fit_cp_2 <- sampling(schools_mod_cp, data = schools_dat,
                     control = list(adapt_delta = 0.999), seed = 978245244)
fit_ncp_2 <- sampling(schools_mod_ncp, data = schools_dat,
                      control = list(adapt_delta = 0.999), seed = 843256842)
```

```{r echo=FALSE, warning=FALSE}
# On rare ocasions, the fits may not be illustrative. Currently the seed is fixed, but if something in Stan changes and the fixed seeds produce unexpected results (which sould be rare), we want to know.
n_divergent_cp_2 <- fit_cp_2 %>% nuts_params() %>% filter(Parameter == "divergent__" & Value == 1) %>% nrow()
n_divergent_ncp_2 <- fit_ncp_2 %>% nuts_params() %>% filter(Parameter == "divergent__" & Value == 1) %>% nrow()

if(n_divergent_cp_2 <= 0) {
  stop("No divergences in CP with increased adapt.delta. Change seed?")
}
if(n_divergent_ncp_2 > 0) {
  stop("Divergences in the NCP model. Fix a bug / change seed?")
}

```

For the first model and this particular data, increasing `adapt_delta` will not
solve the problem and a reparameterization is required.

```{r, mcmc_nuts_divergence-3}
mcmc_nuts_divergence(nuts_params(fit_cp_2), log_posterior(fit_cp_2))
mcmc_nuts_divergence(nuts_params(fit_ncp_2), log_posterior(fit_ncp_2))
```


### Energy and Bayesian fraction of missing information

The `mcmc_nuts_energy` function creates plots similar to those presented in 
Betancourt (2017). While `mcmcm_nuts_divergence` can identify light tails
and incomplete exploration of the target distribution, the `mcmc_nuts_energy` 
function can identify overly heavy tails that are also challenging for sampling.
Informally, the energy diagnostic for HMC (and the related energy-based Bayesian
fraction of missing information) quantifies the heaviness of the tails of the 
posterior distribution.

#### mcmc_nuts_energy

The plot created by `mcmc_nuts_energy` shows overlaid histograms of the 
(centered) marginal energy distribution $\pi_E$ and the first-differenced 
distribution $\pi_{\Delta E}$,
```{r, mcmc_nuts_energy-1, message=FALSE}
color_scheme_set("red")
mcmc_nuts_energy(np_cp)
```

The two histograms ideally look the same (Betancourt, 2017), which is only the
case for the non-centered parameterization (right):

```{r, mcmc_nuts_energy-3, message=FALSE, fig.width=8}
compare_cp_ncp(
  mcmc_nuts_energy(np_cp, binwidth = 1/2),
  mcmc_nuts_energy(np_ncp, binwidth = 1/2)
)
```

The difference between the parameterizations is even more apparent if we force 
the step size to a smaller value and help the chains explore more of the 
posterior:

```{r, mcmc_nuts_energy-4, message=FALSE,  fig.width=8}
np_cp_2 <- nuts_params(fit_cp_2)
np_ncp_2 <- nuts_params(fit_ncp_2)

compare_cp_ncp(
  mcmc_nuts_energy(np_cp_2), 
  mcmc_nuts_energy(np_ncp_2)
)
```

See Betancourt (2017) for more on this particular example as well as the general
theory behind the energy plots.

<br>

## General MCMC diagnostics

A Markov chain generates draws from the target distribution only after it has 
converged to an equilibrium. Unfortunately, this is only guaranteed in the limit in
theory. In practice, diagnostics must be applied to monitor whether the Markov 
chain(s) have converged. The __bayesplot__ package provides various plotting 
functions for visualizing Markov chain Monte Carlo (MCMC) diagnostics after 
fitting a Bayesian model. MCMC draws from any package can be used, although
there are a few diagnostic plots that we will see later in this vignette that
are specifically intended to be used for [Stan](http://mc-stan.org/) models (or
models fit using the same algorithms as Stan).

**Documentation:**

* `help("MCMC-diagnostics")` 
* [mc-stan.org/bayesplot/reference/MCMC-diagnostics](http://mc-stan.org/bayesplot/reference/MCMC-diagnostics.html)

------

### Rhat: potential scale reduction statistic

One way to monitor whether a chain has converged to the equilibrium distribution
is to compare its behavior to other randomly initialized chains. This is the
motivation for the potential scale reduction statistic, split-$\hat{R}$. 
The split-$\hat{R}$ statistic measures the ratio of the average variance of
draws within each chain to the variance of the pooled draws across chains;
if all chains are at equilibrium, these will be the same and $\hat{R}$ will be
one. If the chains have not converged to a common distribution, the $\hat{R}$
statistic will be greater than one (see Gelman et al. 2013, Stan Development Team 2018). 

The **bayesplot** package provides the functions `mcmc_rhat` and
`mcmc_rhat_hist` for visualizing $\hat{R}$ estimates.

First we'll quickly fit one of the models above again, this time intentionally
using too few MCMC iterations and allowing more dispered initial values. 
This should lead to some high $\hat{R}$ values.

```{r, results='hide'}
fit_cp_bad_rhat <- sampling(schools_mod_cp, data = schools_dat, 
                            iter = 50, init_r = 10, seed = 671254821)
```

**bayesplot** provides a generic `rhat` extractor function, currently with
methods defined for models fit using the **rstan**, **rstanarm** and **brms**
packages. But regardless of how you fit your model, all **bayesplot** needs is a
vector of $\hat{R}$ values.

```{r print-rhats}
rhats <- rhat(fit_cp_bad_rhat)
print(rhats)
```

#### mcmc_rhat, mcmc_rhat_hist
We can visualize the $\hat{R}$ values with the `mcmc_rhat` function:

```{r echo=FALSE}
#Check that the fit we got is a sensible example
if(all(rhats < 1.3)) {
  stop("All rhats for the short chain run are low. Change seed?")
}
```


```{r mcmc_rhat-1}
color_scheme_set("brightblue") # see help("color_scheme_set")
mcmc_rhat(rhats)
```

In the plot, the points representing the $\hat{R}$ values are colored based on
whether they are less than $1.05$, between $1.05$ and $1.1$, or greater than
$1.1$. There is no theoretical reason to trichotomize $\hat{R}$ values using
these cutoffs, so keep in mind that this is just a heuristic 

The axis $y$-axis text is off by default for this plot because it's only 
possible to see the labels clearly for models with very few parameters. We can
see the names of the parameters with the concerning $\hat{R}$ values using the
`yaxis_text` convenience function (which passes arguments like `hjust` to
`ggplot2::element_text`):

```{r, mcmc_rhat-2}
mcmc_rhat(rhats) + yaxis_text(hjust = 1)
```

If we look at the same model fit using longer Markov chains we should see all
$\hat{R} < 1.1$, and all points in the plot the same (light) color:

```{r, mcmc_rhat-3}
mcmc_rhat(rhat = rhat(fit_cp)) + yaxis_text(hjust = 0)
```

We can see the same information shown by `mcmc_rhat` but in histogram form using
the `mcmc_rhat_hist` function. See the **Examples** section in 
`help("mcmc_rhat_hist")` for examples.


### Effective sample size

The effective sample size is an estimate of the number of independent draws from
the posterior distribution of the estimand of interest. The $n_{eff}$ metric used in 
Stan is based on the ability of the draws to estimate the true mean value of the
parameter, which is related  to (but not necessarily equivalent to) estimating other
functions of the draws. Because the draws within
a Markov chain are _not_ independent if there is autocorrelation, the effective 
sample size, $n_{eff}$, is usually smaller than the total sample size, $N$ 
(although it may be larger in some cases[^1]). The larger the ratio of $n_{eff}$ to
$N$ the better (see Gelman et al. 2013, Stan Development Team 2018 for more details) .

[^1]: $n_{eff} > N$` indicates that the mean estimate of the parameter computed from Stan 
draws approaches the true mean faster than the mean estimate computed from independent 
samples from the true posterior 
(the estimate from Stan has smaller variance). This is possible when the draws are 
anticorrelated - draws above the mean tend to be well matched with draws below the mean.
Other functions computed from draws (quantiles, posterior intervals, tail probabilities)
may not necessarily approach the true posterior faster. Google "antithetic sampling" or 
visit [the relevant forum thread](http://discourse.mc-stan.org/t/n-eff-bda3-vs-stan/2608/19)
for some further explanation.

The **bayesplot** package provides a generic `neff_ratio` extractor function,
currently with methods defined for models fit using the **rstan**,
**rstanarm** and **brms** packages. But regardless of how you fit your model, all
**bayesplot** needs is a vector of $n_{eff}/N$ values. The `mcmc_neff` and 
`mcmc_neff_hist` can then be used to plot the ratios.

#### mcmc_neff, mcmc_neff_hist

```{r print-neff-ratios}
ratios_cp <- neff_ratio(fit_cp)
print(ratios_cp)
mcmc_neff(ratios_cp, size = 2)
```

In the plot, the points representing the values of $n_{eff}/N$ are colored based
on whether they are less than $0.1$, between $0.1$ and $0.5$, or greater than
$0.5$. These particular values are arbitrary in that they have no particular
theoretical meaning, but a useful heuristic is to worry about any $n_{eff}/N$
less than $0.1$.

One important thing to keep in mind is that these ratios will depend not only on
the model being fit but also on the particular MCMC algorithm used. One reason
why we have such high ratios of $n_{eff}$ to $N$ is that the No-U-Turn sampler
used by **rstan** generally produces draws from the posterior distribution with
much lower autocorrelations compared to draws obtained using other MCMC
algorithms (e.g., Gibbs).

Even for models fit using **rstan** the parameterization can make a big
difference. Here are the $n_{eff}/N$ plots for `fit_cp` and `fit_ncp` 
side by side.

```{r mcmc_neff-compare}
neff_cp <- neff_ratio(fit_cp, pars = c("theta", "mu", "tau"))
neff_ncp <- neff_ratio(fit_ncp, pars = c("theta", "mu", "tau"))
compare_cp_ncp(mcmc_neff(neff_cp), mcmc_neff(neff_ncp), ncol = 1)
```

Because of the difference in parameterization, the effective sample sizes are 
much better for the second model, the non-centered parameterization.


### Autocorrelation

As mentioned above, $n_{eff}/N$ decreases as autocorrelation becomes more 
extreme. We can visualize the autocorrelation using the `mcmc_acf` (line plot) 
or `mcmc_acf_bar` (bar plot) functions. For the selected parameters, these 
functions show the autocorrelation for each Markov chain separately up to a 
user-specified number of lags. Positive autocorrelation is bad (it means the 
chain tends to stay in the same area between iterations) and you want it to 
drop quickly to zero with increasing lag. Negative autocorrelation is possible
and it is useful as it indicates fast convergence of sample mean towards true mean.

#### `mcmc_acf`, `mcmc_acf_bar`

Here we can again see a difference when comparing the two parameterizations of 
the same model. For model 1, $\theta_1$ is the primitive parameter for school 1,
whereas for the non-centered parameterization in model 2 the primitive parameter
is $\eta_1$ (and $\theta_1$ is later constructed from $\eta_1$, $\mu$, 
and $\tau$):

```{r mcmc_acf, out.width = "70%"}
compare_cp_ncp(
  mcmc_acf(posterior_cp, pars = "theta[1]", lags = 10),
  mcmc_acf(posterior_ncp, pars = "eta[1]", lags = 10)
)
```



<br>

## References

Betancourt, M. (2017). A conceptual introduction to Hamiltonian Monte Carlo.
https://arxiv.org/abs/1701.02434

Betancourt, M. (2016). Diagnosing suboptimal cotangent disintegrations in
Hamiltonian Monte Carlo. https://arxiv.org/abs/1604.00695

Betancourt, M. and Girolami, M. (2013). Hamiltonian Monte Carlo for hierarchical
models. https://arxiv.org/abs/1312.0906

Gabry, J., and Goodrich, B. (2018). rstanarm: Bayesian Applied Regression
Modeling via Stan. R package version 2.17.4.
http://mc-stan.org/rstanarm

Gabry, J., Simpson, D., Vehtari, A., Betancourt, M., Gelman, A. (2018).
Visualization in Bayesian workflow. 
_Journal of the Royal Statistical Society Series A_, accepted for publication. 
arXiv preprint: https://arxiv.org/abs/1709.01449

Gelman, A. and Rubin, D. B. (1992). Inference from iterative simulation using
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Gelman, A., Carlin, J. B., Stern, H. S., Dunson, D. B., Vehtari, A., and Rubin,
D. B. (2013). *Bayesian Data Analysis*. Chapman & Hall/CRC Press, London, third
edition.

Hoffman, M. D. and Gelman, A. (2014). The No-U-Turn Sampler: adaptively setting
path lengths in Hamiltonian Monte Carlo. *Journal of Machine Learning Research*.
15:1593--1623.

Rubin, D. B. (1981). Estimation in Parallel Randomized Experiments. *Journal of
Educational and Behavioral Statistics*. 6:377--401.

Stan Development Team. (2018). *Stan Modeling Language Users
Guide and Reference Manual*. http://mc-stan.org/users/documentation/

Stan Development Team. (2018). RStan: the R interface to Stan. R package version 2.17.3.
http://mc-stan.org/rstan
