# hmclearn: Poisson Regression Example

## Introduction

This vignette demonstrates fitting a Poisson regression model via Hamiltonian Monte Carlo (HMC) using the hmclearn package.

library(hmclearn)

For a count response, we let

$p(y | \mu) = \frac{e^{-\mu}\mu^y}{y!},$

with a log-link function

\begin{aligned} \boldsymbol\mu &:= E(\mathbf{y} | \mathbf{X}) = e^{\mathbf{X}\boldsymbol\beta}, \\ \log \boldsymbol\mu &= \mathbf{X}\boldsymbol\beta. \end{aligned} The vector of responses is $$\mathbf{y} = (y_1, ..., y_n)^T$$. The covariate values for subject $$i$$ are $$\mathbf{x}_i^T = (x_{i0}, ..., x_{iq})$$ for $$q$$ covariates plus an intercept. We write the full design matrix as $$\mathbf{X} = (\mathbf{x}_1^T, ..., \mathbf{x}_n^T)^T \in \mathbb{R}^{n\times(q+1)}$$ for $$n$$ observations. The regression coefficients are a vector of length $$q + 1$$, $$\boldsymbol\beta = (\beta_0, ..., \beta_q)^T$$.

## Derive log posterior and gradient for HMC

We develop the likelihood

\begin{aligned} f(\mathbf{y} | \boldsymbol\mu) &= \prod_{i=1}^n \frac{e^{-\mu_i}\mu_i^{y_i}}{y_i!}, \\ f(\mathbf{y} | \mathbf{X}, \boldsymbol\beta) &= \prod_{i=1}^n \frac{e^{-e^{\mathbf{x}_i^T\boldsymbol\beta}}e^{y_i\mathbf{x}_i^T\boldsymbol\beta}}{y_i!}, \\ \end{aligned}

and log-likelihood

\begin{aligned} f(\mathbf{y} | \mathbf{X}, \boldsymbol\beta) &= \sum_{i=1}^n -e^{\mathbf{x}_i^T\boldsymbol\beta} + y_i \mathbf{x}_i^T \boldsymbol\beta - \log y_i!, \\ &\propto \sum_{i=1}^n -e^{\mathbf{x}_i^T\boldsymbol\beta} + y_i \mathbf{x}_i^T \boldsymbol\beta. \end{aligned}

We set a multivariate normal prior for $$\boldsymbol\beta$$

\begin{aligned} \boldsymbol\beta &\sim N(0, \sigma_\beta^2 \mathbf{I}), \end{aligned}

with pdf, omitting constants

\begin{aligned} \pi(\boldsymbol\beta | \sigma_\beta^2) &= \frac{1}{\sqrt{\lvert 2\pi \sigma_\beta^2 \rvert }}e^{-\frac{1}{2}\boldsymbol\beta^T \boldsymbol\beta / \sigma_\beta^2}, \\ \log \pi(\boldsymbol\beta | \sigma_\beta^2) &= -\frac{1}{2}\log(2\pi \sigma_\beta^2) - \frac{1}{2}\boldsymbol\beta^T \boldsymbol\beta / \sigma_\beta^2, \\ &\propto -\frac{1}{2}\log \sigma_\beta^2 - \frac{\boldsymbol\beta^T\boldsymbol\beta}{2\sigma_\beta^2}. \end{aligned}

Next, we derive the log posterior, omitting constants,

\begin{aligned} f(\boldsymbol\beta | \mathbf{X}, \mathbf{y}, \sigma_\beta^2) &\propto f(\mathbf{y} | \mathbf{X}, \boldsymbol\beta) \pi(\boldsymbol\beta | \sigma_\beta^2), \\ \log f(\boldsymbol\beta | \mathbf{X}, \mathbf{y}, \sigma_\beta^2) & \propto \log f(\mathbf{y} | \mathbf{X}, \boldsymbol\beta) + \log \pi(\beta | \sigma_\beta^2), \\ &\propto \sum_{i=1}^n \left( -e^{\mathbf{x}_i^T\boldsymbol\beta} + y_i \mathbf{x}_i^T \boldsymbol\beta\right) - \frac{1}{2}\beta^T\beta/\sigma_\beta^2, \\ &\propto \sum_{i=1}^n \left( -e^{\mathbf{x}_i^T\boldsymbol\beta} + y_i \mathbf{x}_i^T \boldsymbol\beta\right) - \frac{\boldsymbol\beta^T\boldsymbol\beta}{2\sigma_\beta^2}, \\ &\propto \mathbf{y}^T\mathbf{X}\boldsymbol\beta - \mathbf{1}_n^T e^{\mathbf{X}\boldsymbol\beta} - \frac{\boldsymbol\beta^T\boldsymbol\beta}{2\sigma_\beta^2}. \end{aligned}

We need to derive the gradient of the log posterior for the leapfrog function in HMC.

\begin{aligned} \nabla_{\boldsymbol\beta}\log f(\boldsymbol\beta|\mathbf{X}, \mathbf{y}, \sigma_\beta^2) &\propto \sum_{i=1}^n\left( -e^{\mathbf{x}_i^T\boldsymbol\beta}\mathbf{x}_i + y_i\mathbf{x}_i\right) - \boldsymbol\beta / \sigma_\beta^2, \\ &\propto \sum_{i=1}^n \mathbf{x}_i\left( -e^{\mathbf{x}_i^T\boldsymbol\beta} + y_i\right) - \boldsymbol\beta / \sigma_\beta^2, \\ &\propto \mathbf{X}^T (\mathbf{y} - e^{\mathbf{X}\boldsymbol\beta}) - \boldsymbol\beta/ \sigma_\beta^2. \end{aligned}

## Poisson Regression Example Data

The user must define provide the design matrix directly for use in hmclearn. Our first step is to load the data and store the design matrix $$X$$ and dependent variable vector $$y$$.

We load drug usage data and create the design matrix $$X$$ and dependent vector $$y$$. This example also appears in Agresti (2015), and we compare results to his.

data(Drugs)

# design matrix
X <- model.matrix(count ~ A + C + M + A:C + A:M + C:M , data=Drugs)
X <- X[, 1:ncol(X)]

# independent variable is count data
y <- Drugscount ## Comparison model - Frequentist To compare results, we first fit a standard linear model using the frequentist function glm. # matrix representation f <- glm(y ~ X-1, family=poisson(link=log)) summary(f) #> #> Call: #> glm(formula = y ~ X - 1, family = poisson(link = log)) #> #> Deviance Residuals: #> 1 2 3 4 5 6 #> 0.02044 -0.02658 -0.09256 0.02890 -0.33428 0.09452 #> 7 8 #> 0.49134 -0.03690 #> #> Coefficients: #> Estimate Std. Error z value Pr(>|z|) #> X(Intercept) 5.63342 0.05970 94.361 < 2e-16 *** #> XAyes 0.48772 0.07577 6.437 1.22e-10 *** #> XCyes -1.88667 0.16270 -11.596 < 2e-16 *** #> XMyes -5.30904 0.47520 -11.172 < 2e-16 *** #> XAyes:Cyes 2.05453 0.17406 11.803 < 2e-16 *** #> XAyes:Myes 2.98601 0.46468 6.426 1.31e-10 *** #> XCyes:Myes 2.84789 0.16384 17.382 < 2e-16 *** #> --- #> Signif. codes: 0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1 #> #> (Dispersion parameter for poisson family taken to be 1) #> #> Null deviance: 2.4038e+04 on 8 degrees of freedom #> Residual deviance: 3.7399e-01 on 1 degrees of freedom #> AIC: 63.417 #> #> Number of Fisher Scoring iterations: 4 ## Fit model using hmc Next, we fit the poisson regression model using HMC. A vector of $$\epsilon$$ values are specified to align with the data. N <- 1e4 eps_vals <- c(rep(5e-4, 2), 1e-3, 2e-3, 1e-3, 2e-3, 5e-4) set.seed(412) t1.hmc <- Sys.time() f_hmc <- hmc(N = N, theta.init = rep(0, 7), epsilon = eps_vals, L = 50, logPOSTERIOR = poisson_posterior, glogPOSTERIOR = g_poisson_posterior, varnames = colnames(X), parallel=FALSE, chains=2, param=list(y=y, X=X)) t2.hmc <- Sys.time() t2.hmc - t1.hmc #> Time difference of 51.99019 secs ## MCMC summary and diagnostics The acceptance ratio for each of the HMC chains is sufficiently high for an efficient simulation. f_hmcaccept/N
#> [1] 0.9804 0.9829

The posterior quantiles are summarized after removing an initial burnin period. The $$\hat{R}$$ statistics provide an indication of convergence. Values close to one indicate that the multiple MCMC chains coverged to the same distribution, while values above 1.1 indicate possible convergence problems. All $$\hat{R}$$ values in our example are close to one.

summary(f_hmc, burnin=3000, probs=c(0.025, 0.5, 0.975))
#> Summary of MCMC simulation
#>                   2.5%        50%      97.5%     rhat
#> (Intercept)  5.5238199  5.6327435  5.7521131 1.001527
#> Ayes         0.3379506  0.4875867  0.6251244 1.006951
#> Cyes        -2.2176007 -1.8992300 -1.6078705 1.001525
#> Myes        -6.3840770 -5.3643129 -4.4945189 1.007716
#> Ayes:Cyes    1.7495447  2.0670083  2.4061751 1.005151
#> Ayes:Myes    2.1572253  3.0200944  4.1225859 1.018774
#> Cyes:Myes    2.5174705  2.8574315  3.1983475 1.026709

Trace plots provide a visual indication of stationarity. These plots indicate that the MCMC chains are reasonably stationary.

mcmc_trace(f_hmc, burnin=3000)

Histograms of the posterior distribution show that Bayesian parameter estimates align with Frequentist estimates.

beta.freq <- coef(f)
diagplots(f_hmc, burnin=3000, comparison.theta=beta.freq)
#> \$histogram

## Source

Data originally provided by Harry Khamis, Wright State University

## References

Agresti, A. (2015). Foundations of linear and generalized linear models. John Wiley & Sons. ISBN: 978-1-118-73003-4

Thomas, Samuel, and Wanzhu Tu. “Learning Hamiltonian Monte Carlo in R.” arXiv preprint arXiv:2006.16194 (2020).