Author: Topi Paananen, Paul Bürkner, Aki Vehtari and Jonah Gabry
Date: 2024-07-03

# Avoiding model refits in leave-one-out cross-validation with moment matching

#### Topi Paananen, Paul Bürkner, Aki Vehtari and Jonah Gabry

#### 2024-07-03

# Introduction

This vignette demonstrates how to improve the Monte Carlo sampling accuracy of leave-one-out cross-validation with the loo package and Stan. The loo package automatically monitors the sampling accuracy using Pareto \(k\) diagnostics for each observation. Here, we present a method for quickly improving the accuracy when the Pareto diagnostics indicate problems. This is done by performing some additional computations using the existing posterior sample. If successful, this will decrease the Pareto \(k\) values, making the model assessment more reliable. loo also stores the original Pareto \(k\) values with the name `influence_pareto_k` which are not changed. They can be used as a diagnostic of how much each observation influences the posterior distribution.

The methodology presented is based on the paper

- Paananen, T., Piironen, J., Buerkner, P.-C., Vehtari, A. (2020). Implicitly Adaptive Importance Sampling. arXiv preprint arXiv:1906.08850.

More information about the Pareto \(k\) diagnostics is given in the following papers

- Vehtari, A., Gelman, A., and Gabry, J. (2017). Practical Bayesian model evaluation using leave-one-out cross-validation and WAIC. Statistics and Computing. 27(5), 1413–1432. :10.1007/s11222-016-9696-4. Links: published | arXiv preprint.

- Vehtari, A., Simpson, D., Gelman, A., Yao, Y., and Gabry, J. (2024). Pareto smoothed importance sampling. Journal of Machine Learning Research, 25(72):1-58. PDF

# Example: Eradication of Roaches

We will use the same example as in the vignette Using the loo package (version >= 2.0.0). See the demo for a description of the problem and data. We will use the same Poisson regression model as in the case study.

## Coding the Stan model

Here is the Stan code for fitting the Poisson regression model, which we will use for modeling the number of roaches.

    # Note: some syntax used in this Stan program requires RStan >= 2.26 (or CmdStanR)
    # To use an older version of RStan change the line declaring `y` to: int y[N];
    stancode <- "
    data {
      int<lower=1> K;
      int<lower=1> N;
      matrix[N,K] x;
      array[N] int y;
      vector[N] offset;

      real beta_prior_scale;
      real alpha_prior_scale;
    }
    parameters {
      vector[K] beta;
      real intercept;
    }
    model {
      y ~ poisson(exp(x * beta + intercept + offset));
      beta ~ normal(0,beta_prior_scale);
      intercept ~ normal(0,alpha_prior_scale);
    }
    generated quantities {
      vector[N] log_lik;
      for (n in 1:N)
        log_lik[n] = poisson_lpmf(y[n] | exp(x[n] * beta + intercept + offset[n]));
    }
    "

Following the usual approach recommended in Writing Stan programs for use with the loo package, we compute the log-likelihood for each observation in the `generated quantities` block of the Stan program.

## Setup

In addition to loo, we load the rstan package for fitting the model, and the rstanarm package for the data.

    library("rstan")
    library("loo")
    seed <- 9547
    set.seed(seed)

## Fitting the model with RStan

Next we fit the model in Stan using the rstan package:

    # Prepare data
    data(roaches, package = "rstanarm")
    roaches$roach1 <- sqrt(roaches$roach1)
    y <- roaches$y
    x <- roaches[,c("roach1", "treatment", "senior")]
    offset <- log(roaches[,"exposure2"])
    n <- dim(x)[1]
    k <- dim(x)[2]

    standata <- list(N = n, K = k, x = as.matrix(x), y = y, offset = offset, beta_prior_scale = 2.5, alpha_prior_scale = 5.0)

    # Compile
    stanmodel <- stan_model(model_code = stancode)

    # Fit model
    fit <- sampling(stanmodel, data = standata, seed = seed, refresh = 0)
    print(fit, pars = "beta")

    Inference for Stan model: anon_model.
    4 chains, each with iter=2000; warmup=1000; thin=1;
    post-warmup draws per chain=1000, total post-warmup draws=4000.

             mean se_mean   sd  2.5%   25%   50%   75% 97.5% n_eff Rhat
    beta[1]  0.16       0 0.00  0.16  0.16  0.16  0.16  0.16  2344    1
    beta[2] -0.57       0 0.03 -0.62 -0.59 -0.57 -0.55 -0.52  2395    1
    beta[3] -0.31       0 0.03 -0.38 -0.34 -0.31 -0.29 -0.25  2135    1

    Samples were drawn using NUTS(diag_e) at Wed Jul  3 10:22:19 2024.
    For each parameter, n_eff is a crude measure of effective sample size,
    and Rhat is the potential scale reduction factor on split chains (at
    convergence, Rhat=1).

Let us now evaluate the predictive performance of the model using `loo()`.

    loo1 <- loo(fit)

    Warning: Some Pareto k diagnostic values are too high. See help('pareto-k-diagnostic') for details.

    loo1

    Computed from 4000 by 262 log-likelihood matrix.

             Estimate     SE
    elpd_loo  -5459.4  694.1
    p_loo       258.8   55.4
    looic     10918.9 1388.2
    ------
    MCSE of elpd_loo is NA.
    MCSE and ESS estimates assume MCMC draws (r_eff in [0.4, 1.0]).

    Pareto k diagnostic values:
                             Count Pct.    Min. ESS
    (-Inf, 0.7]   (good)     248   94.7%   53
       (0.7, 1]   (bad)        7    2.7%   <NA>
       (1, Inf)   (very bad)   7    2.7%   <NA>
    See help('pareto-k-diagnostic') for details.

The `loo()` function output warnings that there are some observations which are highly influential, and thus the accuracy of importance sampling is compromised as indicated by the large Pareto \(k\) diagnostic values (> 0.7). As discussed in the vignette Using the loo package (version >= 2.0.0), this may be an indication of model misspecification. Despite that, it is still beneficial to be able to evaluate the predictive performance of the model accurately.

## Moment matching correction for importance sampling

To improve the accuracy of the `loo()` result above, we could perform leave-one-out cross-validation by explicitly leaving out single observations and refitting the model using MCMC repeatedly. However, the Pareto \(k\) diagnostics indicate that there are 19 observations which are problematic. This would require 19 model refits which may require a lot of computation time.

Instead of refitting with MCMC, we can perform a faster moment matching correction to the importance sampling for the problematic observations. This can be done with the `loo_moment_match()` function in the loo package, which takes our existing `loo` object as input and modifies it. The moment matching requires some evaluations of the model posterior density. For models fitted with rstan, this can be conveniently done by using the existing `stanfit` object.

First, we show how the moment matching can be used for a model fitted using rstan. It only requires setting the argument `moment_match` to `TRUE` in the `loo()` function. Optionally, you can also set the argument `k_threshold` which determines the Pareto \(k\) threshold, above which moment matching is used. By default, it operates on all observations whose Pareto \(k\) value is larger than the sample size (\(S\)) specific threshold \(\min(1 - 1 / \log_{10}(S), 0.7)\) (which is \(0.7\) for \(S>2200\)).

    # available in rstan >= 2.21
    loo2 <- loo(fit, moment_match = TRUE)
    loo2

    Computed from 4000 by 262 log-likelihood matrix.

             Estimate     SE
    elpd_loo  -5478.8  700.0
    p_loo       278.2   63.2
    looic     10957.7 1400.1
    ------
    MCSE of elpd_loo is 0.4.
    MCSE and ESS estimates assume MCMC draws (r_eff in [0.4, 1.0]).

    All Pareto k estimates are good (k < 0.7).
    See help('pareto-k-diagnostic') for details.

After the moment matching, all observations have the diagnostic Pareto \(k\) less than 0.7, meaning that the estimates are now reliable. The total `elpd_loo` estimate also changed from `-5457.8` to `-5478.5`, showing that before moment matching, `loo()` overestimated the predictive performance of the model.

The updated Pareto \(k\) values stored in `loo2$diagnostics$pareto_k` are considered algorithmic diagnostic values that indicate the sampling accuracy. The original Pareto \(k\) values are stored in `loo2$pointwise[,"influence_pareto_k"]` and these are not modified by the moment matching. These can be considered as diagnostics for how big influence each observation has on the posterior distribution. In addition to the Pareto \(k\) diagnostics, moment matching also updates the effective sample size estimates.

# Using `loo_moment_match()` directly

The moment matching can also be performed by explicitly calling the function `loo_moment_match()`. This enables its use also for models that are not using rstan or another package with built-in support for `loo_moment_match()`. To use `loo_moment_match()`, the user must give the model object `x`, the `loo` object, and 5 helper functions as arguments to `loo_moment_match()`. The helper functions are

- `post_draws`
  - A function the takes `x` as the first argument and returns a matrix of posterior draws of the model parameters, `pars`.
- `log_lik_i`
  - A function that takes `x` and `i` and returns a matrix (one column per chain) or a vector (all chains stacked) of log-likeliood draws of the ith observation based on the model `x`. If the draws are obtained using MCMC, the matrix with MCMC chains separated is preferred.
- `unconstrain_pars`
  - A function that takes arguments `x` and `pars`, and returns posterior draws on the unconstrained space based on the posterior draws on the constrained space passed via `pars`.
- `log_prob_upars`
  - A function that takes arguments `x` and `upars`, and returns a matrix of log-posterior density values of the unconstrained posterior draws passed via `upars`.
- `log_lik_i_upars`
  - A function that takes arguments `x`, `upars`, and `i` and returns a vector of log-likelihood draws of the `i`th observation based on the unconstrained posterior draws passed via `upars`.

Next, we show how the helper functions look like for RStan objects, and show an example of using `loo_moment_match()` directly. For stanfit objects from rstan objects, the functions look like this:

    # create a named list of draws for use with rstan methods
    .rstan_relist <- function(x, skeleton) {
      out <- utils::relist(x, skeleton)
      for (i in seq_along(skeleton)) {
        dim(out[[i]]) <- dim(skeleton[[i]])
      }
      out
    }

    # rstan helper function to get dims of parameters right
    .create_skeleton <- function(pars, dims) {
      out <- lapply(seq_along(pars), function(i) {
        len_dims <- length(dims[[i]])
        if (len_dims < 1) return(0)
        return(array(0, dim = dims[[i]]))
      })
      names(out) <- pars
      out
    }

    # extract original posterior draws
    post_draws_stanfit <- function(x, ...) {
      as.matrix(x)
    }

    # compute a matrix of log-likelihood values for the ith observation
    # matrix contains information about the number of MCMC chains
    log_lik_i_stanfit <- function(x, i, parameter_name = "log_lik", ...) {
      loo::extract_log_lik(x, parameter_name, merge_chains = FALSE)[, , i]
    }

    # transform parameters to the unconstraint space
    unconstrain_pars_stanfit <- function(x, pars, ...) {
      skeleton <- .create_skeleton(x@sim$pars_oi, x@par_dims[x@sim$pars_oi])
      upars <- apply(pars, 1, FUN = function(theta) {
        rstan::unconstrain_pars(x, .rstan_relist(theta, skeleton))
      })
      # for one parameter models
      if (is.null(dim(upars))) {
        dim(upars) <- c(1, length(upars))
      }
      t(upars)
    }

    # compute log_prob for each posterior draws on the unconstrained space
    log_prob_upars_stanfit <- function(x, upars, ...) {
      apply(upars, 1, rstan::log_prob, object = x,
            adjust_transform = TRUE, gradient = FALSE)
    }

    # compute log_lik values based on the unconstrained parameters
    log_lik_i_upars_stanfit <- function(x, upars, i, parameter_name = "log_lik",
                                      ...) {
      S <- nrow(upars)
      out <- numeric(S)
      for (s in seq_len(S)) {
        out[s] <- rstan::constrain_pars(x, upars = upars[s, ])[[parameter_name]][i]
      }
      out
    }

Using these function, we can call `loo_moment_match()` to update the existing `loo` object.

    loo3 <- loo::loo_moment_match.default(
      x = fit,
      loo = loo1,
      post_draws = post_draws_stanfit,
      log_lik_i = log_lik_i_stanfit,
      unconstrain_pars = unconstrain_pars_stanfit,
      log_prob_upars = log_prob_upars_stanfit,
      log_lik_i_upars = log_lik_i_upars_stanfit
    )
    loo3

    Computed from 4000 by 262 log-likelihood matrix.

             Estimate     SE
    elpd_loo  -5478.8  700.0
    p_loo       278.2   63.2
    looic     10957.7 1400.1
    ------
    MCSE of elpd_loo is 0.4.
    MCSE and ESS estimates assume MCMC draws (r_eff in [0.4, 1.0]).

    All Pareto k estimates are good (k < 0.7).
    See help('pareto-k-diagnostic') for details.

As expected, the result is identical to the previous result of `loo2 <- loo(fit, moment_match = TRUE)`.

# References

Gelman, A., and Hill, J. (2007). Data Analysis Using Regression and Multilevel Hierarchical Models. Cambridge University Press.

Stan Development Team (2020) RStan: the R interface to Stan, Version 2.21.1 https://mc-stan.org

Paananen, T., Piironen, J., Buerkner, P.-C., Vehtari, A. (2021). Implicitly adaptive importance sampling. Statistics and Computing, 31, 16. :10.1007/s11222-020-09982-2. arXiv preprint arXiv:1906.08850.

Vehtari, A., Gelman, A., and Gabry, J. (2017). Practical Bayesian model evaluation using leave-one-out cross-validation and WAIC. Statistics and Computing. 27(5), 1413–1432. :10.1007/s11222-016-9696-4. Links: published | arXiv preprint.

Vehtari, A., Simpson, D., Gelman, A., Yao, Y., and Gabry, J. (2024). Pareto smoothed importance sampling. Journal of Machine Learning Research, 25(72):1-58. PDF