# Chapter 12 Poisson and Negative Binomial regression

Step back from the details of the previous few chapters and recall the big goal: to build **regression** models of **quantitative** response variables \(Y\).
We’ve only shared one regression tool with you so far, the *Bayesian Normal regression model*.
The name of this “Normal” regression tool reflects its broad applicability.
But (luckily!), not *every* model is “Normal.”
We’ll expand upon our regression tools in the context of the following data story.

As of this book’s writing, the *Equality Act* sits in the United States Senate awaiting consideration.
If passed, this act or bill would ensure basic LGBTQ+ rights at the *national* level by prohibiting discrimination in education, employment, housing, and more.
As is, each of the 50 individual states has their own set of unique anti-discrimination laws, spanning issues from anti-bullying to health care coverage.
Our goal is to better understand how the *number* of laws in a state relates to its unique demographic features and political climate.
For the former, we’ll narrow our focus to the percentage of a state’s residents that reside in an urban area.
For the latter, we’ll utilize historical voting patterns in presidential elections, noting whether a state has consistently voted for the *Democratic* candidate, consistently voted for the “GOP” Republican candidate^{58}, or is a *swing* state that has flip flopped back and forth.
Throughout our analysis, please recognize that the *number* of laws is *not* a perfect proxy for the *quality* of a state’s laws – it merely provides a starting point in understanding how laws vary from state to state.

For each state \(i \in \{1,2,\ldots,50\}\), let \(Y_i\) denote the number of anti-discrimination laws and predictor \(X_{i1}\) denote the percentage of the state’s residents that live in urban areas.
Further, our historical political climate predictor variable is categorical with three levels: Democrat, GOP, or swing.
This is our first time working with a three level variable, so let’s set this up right.
Recall from Chapter 11 that one level of a categorical predictor, here Democrat, serves as a **baseline** or **reference level** for our model.
The *other* levels, GOP and swing, enter our model as *indicators*.
Thus \(X_{i2}\) indicates whether state \(i\) leans GOP and \(X_{i3}\) indicates a swing state:

\[X_{i2} = \begin{cases} 1 & \text{ GOP} \\ 0 & \text{ otherwise} \\ \end{cases} \;\;\;\; \text{ and } \;\;\;\; X_{i3} = \begin{cases} 1 & \text{ swing} \\ 0 & \text{ otherwise} \\ \end{cases}\]

Since it’s the only technique we’ve explored thus far, our first approach to *understanding* the relationship between our **quantitative** response variable \(Y\) and our predictors \(X\) might be to build a regression model with a Normal data structure:

\[Y_i | \beta_0, \beta_1, \beta_2, \beta_3, \sigma \stackrel{ind}{\sim} N\left(\mu_i, \; \sigma^2 \right) \;\; \text{ with } \;\; \mu_i = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \beta_3 X_{i3}.\]

Other than an understanding that a state that’s “typical” with respect to its urban population and historical voting patterns has around 7 laws, we have very little prior knowledge about this relationship. Thus we’ll set a \(N(7, 1.5^2)\) prior for the centered intercept \(\beta_{0c}\), but utilize weakly informative priors for the other three parameters.

Next, let’s consider some data.
Each year, the Human Rights Campaign releases a “State Equality Index” which monitors the *number* of LQBTQ+ rights laws in each state.
Among other state features, the `equality_index`

dataset in the `bayesrules`

package includes data from the 2019 index compiled by Sarah Warbelow, Courtnay Avant, and Colin Kutney (2019).
To obtain a detailed codebook, type `?equality`

in the console.

```
# Load packages
library(bayesrules)
library(rstanarm)
library(bayesplot)
library(tidyverse)
library(tidybayes)
library(broom.mixed)
# Load data
data(equality_index)
<- equality_index equality
```

The histogram below indicates that the number of laws ranges from as low as 1 to as high as 155, yet the majority of states have fewer than ten laws:

```
ggplot(equality, aes(x = laws)) +
geom_histogram(color = "white", breaks = seq(0, 160, by = 10))
```

The state with 155 laws happens to be California. As a clear outlier, we’ll remove this state from our analysis:

```
# Identify the outlier
%>%
equality filter(laws == max(laws))
# A tibble: 1 x 6
state region gop_2016 laws historical percent_urban<fct> <fct> <dbl> <dbl> <fct> <dbl>
1 calif… west 31.6 155 dem 95
# Remove the outlier
<- equality %>%
equality filter(state != "california")
```

Next, in a scatterplot of the number of state `laws`

versus its `percent_urban`

population and `historical`

voting patterns, notice that historically `dem`

states and states with greater urban populations tend to have more LGBTQ+ anti-discrimination laws in place:

```
ggplot(equality, aes(y = laws, x = percent_urban, color = historical)) +
geom_point()
```

Using `stan_glm()`

, we combine this data with our weak prior understanding to simulate the posterior Normal regression model of `laws`

by `percent_urban`

and `historical`

voting trends.
In a quick posterior predictive check of this `equality_normal_sim`

model, we compare a histogram of the observed anti-discrimination laws to five posterior simulated data sets (Figure 12.3).
(A histogram is more appropriate than a density plot here since our response variable is a non-negative count.)
The results aren’t good – the posterior predictions from this model do *not* match the features of the observed data.
You might not be surprised.
The observed number of anti-discrimination laws per state are right skewed (not Normal!).
In contrast, the data sets simulated from the posterior Normal regression model are roughly symmetric.
Adding insult to injury, these simulated data sets assume that it is quite common for states to have a *negative* number of laws (not possible!).

```
# Simulate the Normal model
<- stan_glm(laws ~ percent_urban + historical,
equality_normal_sim data = equality,
family = gaussian,
prior_intercept = normal(7, 1.5),
prior = normal(0, 2.5, autoscale = TRUE),
prior_aux = exponential(1, autoscale = TRUE),
chains = 4, iter = 5000*2, seed = 84735)
# Posterior predictive check
pp_check(equality_normal_sim, plotfun = "hist", nreps = 5) +
geom_vline(xintercept = 0) +
xlab("laws")
```

This sad result reveals the limits of our lone tool. Luckily, probability models come in all shapes, and we don’t have to force something to be Normal when it’s not.

You will extend the Normal regression model of a quantitative response variable \(Y\) to settings in which \(Y\) is a **count** variable whose dependence on predictors \(X\) is better represented by the **Poisson** or **Negative Binomial**, not Normal, models.

## 12.1 Building the Poisson regression model

### 12.1.1 Specifying the data model

Recall from Chapter 5 that the Poisson model is appropriate for modeling discrete counts of events (here anti-discrimination laws) that happen in a fixed interval of space or time (here states) and that, *theoretically*, have no upper bound.
The Poisson is especially handy in cases like ours in which counts are right-skewed, thus can’t reasonably be approximated by a Normal model.
Moving forward, let’s assume a **Poisson data model** for the number of LGBTQ+ anti-discrimination laws in each state \(i\) (\(Y_i\)) where the *rate* of anti-discrimination laws \(\lambda_i\) depends upon demographic and voting trends (\(X_{i1}\), \(X_{i2}\), and \(X_{i3}\)):

\[Y_i | \lambda_i \stackrel{ind}{\sim} Pois\left(\lambda_i \right) \; .\]

Under this Poisson structure, the *expected* or *average* number of laws \(Y_i\) in states with similar predictor values \(X\) is captured by \(\lambda_i\):

\[E(Y_i | \lambda_i) = \lambda_i \; . \]

If we proceeded as in Normal regression, we might assume that the average number of laws is a linear combination of our predictors, \(\lambda_i = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \beta_3 X_{i3}\). This assumption is illustrated by the lines in Figure 12.4, one per historical voting trend.

Figure 12.4 highlights a flaw in assuming that the expected number of laws in a state, \(\lambda_i\), is a linear combination of `percent_urban`

and `historical`

. What is it?

When we assume that \(\lambda_i\) can be expressed by a linear combination of the \(X\) predictors, the model of \(\lambda_i\) spans both positive and negative values, thus suggests that some states have a *negative* number of anti-discrimination laws.
That doesn’t make sense.
Like the number of laws, a Poisson *rate* \(\lambda_i\), must be *positive*.
To avoid this violation, it is common to use a **log link function**.^{59}
That is, we’ll assume that \(log(\lambda_i)\), which *does* span both positive and negative values, is a linear combination of the \(X\) predictors:

\[Y_i | \beta_0,\beta_1, \beta_2, \beta_3 \stackrel{ind}{\sim} Pois\left(\lambda_i \right) \;\;\; \text{ with } \;\;\; \log\left( \lambda_i \right) = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \beta_3 X_{i3} \; .\]

At the risk of projecting, interpreting the *logged* number of laws isn’t so easy.
Instead, we can always transform the model relationship off the log scale by appealing to the fact that if \(\log(\lambda) = a\), then \(\lambda = e^a\) for natural number \(e\):

\[Y_i | \beta_0,\beta_1, \beta_2, \beta_3 \stackrel{ind}{\sim} Pois\left(\lambda_i\right) \;\; \text{ with } \;\; \lambda_i = e^{\beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \beta_3 X_{i3}} \; .\]

Figure 12.5 presents a prior plausible outcome for this model on both the \(\log(\lambda)\) and \(\lambda\) scales.
In both cases there are three curves, one per historical voting category.
On the \(\log(\lambda)\) scale, these curves are *linear*.
Yet when transformed to the \(\lambda\) or (unlogged) laws scale, these curves are *nonlinear* and restricted to be at or above 0.
This was the point – we *want* our model to preserve the fact that a state can’t have a negative number of laws.

The model curves on both the \(\log(\lambda)\) and \(\lambda\) scales are defined by the (\(\beta_0, \beta_1, \beta_2, \beta_3)\) parameters.
When describing the *linear* model of the *logged* number of laws in a state, these parameters take on the usual meanings related to intercepts and slopes.
Yet (\(\beta_0, \beta_1, \beta_2, \beta_3)\) take on new meanings when describing the *nonlinear* model of the (unlogged) number of laws in a state.

**Interpreting Poisson regression coefficients**

Consider two equivalent formulations of the Poisson regression model: \(Y|\beta_0,\beta_1,\ldots,\beta_p \sim \text{Pois}(\lambda)\) with

\[\begin{equation} \begin{split} \log(\lambda) & = \beta_0 + \beta_1 X_1 + \cdots \beta_p X_p \\ \lambda & = e^{\beta_0 + \beta_1 X_1 + \cdots \beta_p X_p} \\ \end{split} \tag{12.1} \end{equation}\]

**Interpreting \(\beta_0\)**

When \(X_1, X_2, \ldots, X_p\) are all 0, \(\beta_0\) is the *logged* average \(Y\) value and \(e^{\beta_0}\) is the average \(Y\) value.

**Interpreting \(\beta_1\)** (and similarly \(\beta_2, \ldots, \beta_p\))

Let \(\lambda_x\) represent the average \(Y\) value when \(X_1 = x\) and \(\lambda_{x+1}\) represent the average \(Y\) value when \(X_1 = x + 1\).
When we control for all other predictors \(X_2, \ldots, X_p\) and increase \(X_1\) by 1, from \(x\) to \(x + 1\): \(\beta_1\) is the change in the *logged* average \(Y\) value and \(e^{\beta_1}\) is the *multiplicative change* in the (unlogged) average \(Y\) value.
That is,

\[\beta_1 = \log(\lambda_{x+1}) - \log(\lambda_x) \;\;\;\; \text{ and } \;\;\;\; e^{\beta_1} = \frac{\lambda_{x+1}}{\lambda_x} \; .\]

Let’s apply these concepts to the hypothetical models of the logged and unlogged number of state laws in Figure 12.5. In this figure, the curves are defined by:

\[\begin{equation} \begin{split} \log(\lambda) & = 2 + 0.03X_{i1} - 1.1X_{i2} - 0.6X_{i3} \\ \lambda & = e^{2 + 0.03X_{i1} - 1.1X_{i2} - 0.6X_{i3}} \\ \end{split} \tag{12.2} \end{equation}\]

Consider the intercept parameter \(\beta_0 = 2\).
This intercept reflects the trends in anti-discrimination laws in historically Democratic states with 0 urban population, i.e. states with \(X_{i1} = X_{i2} = X_{i3} = 0\).
As such, we expect the *logged* number of laws in such states to be \(\beta_0 = 2\).
More meaningfully, and equivalently, we expect historically Democratic states with 0 urban population to have roughly 7.4 anti-discrimination laws:

\[e^{\beta_0} = e^2 = 7.389 \;.\]

Now, since there *are* no states in which the urban population is close to 0, it doesn’t make sense to put too much emphasis on these interpretations of \(\beta_0\).
Rather, we can just understand \(\beta_0\) as providing a baseline for our model, on both the \(\log(\lambda)\) and \(\lambda\) scales.

Next, consider the urban percentage coefficient \(\beta_1 = 0.03\).
On the *linear* \(\log(\lambda)\) scale, we can still interpret this value as the shared *slope* of the lines in Figure 12.5 (left).
Specifically, no matter a state’s `historical`

voting trends, we expect the *logged* number of laws in states to increase by 0.03 for every extra percentage point in urban population.
We can interpret the relationship between state laws and urban percentage more meaningfully on the *unlogged* scale of \(\lambda\) by examining

\[e^{\beta_1} = e^{0.03} = 1.03 \; .\]

To this end, reexamine the *nonlinear* models of anti-discrimination laws in Figure 12.5 (right).
Though the model curves don’t increase by the same *absolute* amount for each incremental increase in `percent_urban`

, they do increase by the same *percentage*.
Thus instead of representing a linear slope, \(e^{\beta_1}\) measures the nonlinear *percentage* or *multiplicative* increase in the average number of laws with urban percentage.
In this case, if the urban population in one state is 1 percentage point greater than another state, we’d expect it to have 1.03 *times* the number of, or 3% more, anti-discrimination laws.

Finally, consider the GOP coefficient \(\beta_2 = -1.1\).
Recall from Chapter 11 that when interpreting a coefficient for a categorical indicator, we must do so relative to the reference category, here Democrat leaning states.
On the *linear* \(\log(\lambda)\) scale, \(\beta_2\) serves as the vertical *difference* between the GOP versus Democrat model lines in Figure 12.5 (left).
Thus at any urban percentage, we’d expect the *logged* number of laws to be 1.1 lower in historically GOP states than Democrat states.
Though we also see a difference between the GOP and Democrat curves on the *nonlinear* \(\lambda\) scale (Figure 12.5 right), the difference isn’t constant – the gap in the number of laws widens as urban percentage increases.
In this case, instead of representing a constant difference between two *lines*, \(e^{\beta_2}\) measures the *percentage* or *multiplicative* difference in the GOP versus Democrat curve.
Thus at any urban percentage level, we’d expect a historically GOP state to have 1/3 as many anti-discrimination laws as a historically Democrat state:

\[e^{\beta_2} = e^{-1.1} = 0.333 \; .\]

We conclude this section with the formal assumptions encoded by the Poisson data model.

**Poisson regression assumptions**

Consider the Bayesian Poisson regression model with \(Y_i | \beta_0,\beta_1, \ldots, \beta_p \stackrel{ind}{\sim} Pois\left(\lambda_i \right)\) where

\[\log\left( \lambda_i \right) = \beta_0 + \beta_1 X_{i1} + \cdots + \beta_p X_{ip} \; .\]

The appropriateness of this model depends upon the following assumptions.

**Structure of the data**

Conditioned on predictors \(X\), the observed data \(Y_i\) on case \(i\) is**independent**of the observed data on any other case \(j\).**Structure of variable \(Y\)**

Response variable \(Y\) has a Poisson structure, i.e. is a discrete**count**of events that happen in a fixed interval of space or time.**Structure of the relationship**

The*logged*average \(Y\) value can be written as a**linear combination**of the predictors, \(\log\left( \lambda_i \right) = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2}+ \beta_3 X_{i3}\).**Structure of the variability in \(Y\)**

A Poisson random variable \(Y\) with rate \(\lambda\) has equal mean and variance, \(E(Y) = \text{Var}(Y) = \lambda\) (5.3). Thus conditioned on predictors \(X\), the*typical*value of \(Y\) should be roughly equivalent to the*variability*in \(Y\). As such, the variability in \(Y\) increases as its mean increases. See Figure 12.6 for examples of when this assumption does and does not hold.

### 12.1.2 Specifying the priors

To complete our Bayesian model, we must express our prior understanding of regression coefficients \((\beta_0, \beta_1, \beta_2, \beta_3)\).
Since these coefficients can each take on any value on the real line, it’s again reasonable to utilize Normal priors.
Further, as was the case for our Normal regression model, we’ll assume these priors (i.e. our prior understanding of the model coefficients) are **independent**.
The complete representation of our **Poisson regression model** of \(Y_i\) is as follows:

\[\begin{equation} \begin{array}{lcrl} \text{data:} & \hspace{.025in} & Y_i|\beta_0,\beta_1,\beta_2,\beta_3 & \stackrel{ind}{\sim} Pois\left(\lambda_i \right) \;\; \text{ with } \;\; \log\left( \lambda_i \right) = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2}+ \beta_3 X_{i3} \\ \text{priors:} & & \beta_{0c} & \sim N\left(2, 0.5^2 \right) \\ && \beta_1 & \sim N(0, 0.17^2) \\ && \beta_2 & \sim N(0, 4.97^2) \\ && \beta_3 & \sim N(0, 5.60^2) \\ \end{array} \tag{12.3} \end{equation}\]

First, consider the prior for the centered intercept \(\beta_{0c}\).
Recall our prior assumption that the average number of laws in “typical” states is around \(\lambda = 7\).
As such, we set the Normal prior mean for \(\beta_{0c}\) to 2 on the *log* scale:

\[\log(\lambda) = \log(7) \approx 1.95 \; .\]

Further, the range of this Normal prior indicates our relative uncertainty about this baseline.
Though the *logged* average number of laws is most likely around 2, we think it could range from roughly 1 to 3 (\(2 \pm 2*0.5\)).
Or, more intuitively, we think that the average *number* of laws in typical states might be somewhere between 3 and 20:

\[(e^1, e^3) \approx (3, 20)\;.\]

Beyond this baseline, we again used weakly informative default priors for (\(\beta_1,\beta_2,\beta_3\)), tuned by `stan_glm()`

below.
Being centered at zero with relatively large standard deviation on the scale of our variables, these priors reflect a general uncertainty about whether and how the number of anti-discrimination laws is associated with a state’s urban population and voting trends.

To examine whether these combined priors accurately reflect our uncertain understanding of state laws, we’ll simulate 20,000 draws from the prior models of \((\beta_0, \beta_1, \beta_2, \beta_3)\).
To this end, we can run the same `stan_glm()`

function that we use to simulate the posterior with two new arguments: `prior_PD = TRUE`

specifies that we wish to simulate the *prior*, and `family = poisson`

indicates that we’re using a Poisson data model (not Normal or `gaussian`

).

```
<- stan_glm(laws ~ percent_urban + historical,
equality_model_prior data = equality,
family = poisson,
prior_intercept = normal(2, 0.5),
prior = normal(0, 2.5, autoscale = TRUE),
chains = 4, iter = 5000*2, seed = 84735,
prior_PD = TRUE)
```

A call to `prior_summary()`

confirms the specification of our weakly informative priors in (12.3):

`prior_summary(equality_model_prior)`

Next, we plot 100 prior plausible models, \(e^{\beta_0 + \beta_1x_1 + \beta_2x_2 + \beta_3x_3}\).
The mess of curves here certainly reflects our prior uncertainty!
These are all over the map, indicating that the number of laws might increase *or* decrease with urban population and might or might not differ by historical voting trends.
We don’t really know.

```
%>%
equality add_fitted_draws(equality_model_prior, n = 100) %>%
ggplot(aes(x = percent_urban, y = laws, color = historical)) +
geom_line(aes(y = .value, group = paste(historical, .draw))) +
ylim(0, 100)
```

## 12.2 Simulating the posterior

With all the pieces in place, let’s simulate the Poisson posterior regression model of anti-discrimination laws versus urban population and voting trends (12.3).
Instead of starting from scratch with the `stan_glm()`

function, we’ll take a shortcut: we `update()`

the `equality_model_prior`

simulation, indicating `prior_PD = FALSE`

(i.e. we wish to simulate the posterior, not the prior).

`<- update(equality_model_prior, prior_PD = FALSE) equality_model `

MCMC trace, density, and autocorrelation plots confirm that our simulation has stabilized:

```
mcmc_trace(equality_model)
mcmc_dens_overlay(equality_model)
mcmc_acf(equality_model)
```

And before we get too far into our analysis of these simulation results, a quick posterior predictive check confirms that we’re now on the right track (Figure 12.8).
First, histograms of just five posterior simulations of state law data exhibit similar skew, range, and trends as the observed law data.
Second, though density plots aren’t the best display of count data, they allow us to more *directly* compare a *broader* range of 50 posterior simulated data sets to the actual observed law data.
These simulations aren’t perfect, but they do reasonably capture the features of the observed law data.

```
set.seed(1)
pp_check(equality_model, plotfun = "hist", nreps = 5) +
xlab("laws")
pp_check(equality_model) +
xlab("laws")
```

## 12.3 Interpreting the posterior

With the reassurance that our model isn’t too wrong, let’s dig into the posterior results, beginning with the big picture. The 50 posterior plausible models in Figure 12.9, \(\lambda = e^{\beta_0 + \beta_1X_1 + \beta_2X_2 + \beta_3X_3}\), provide insight into our overall understanding of anti-discrimination laws.

```
%>%
equality add_fitted_draws(equality_model, n = 50) %>%
ggplot(aes(x = percent_urban, y = laws, color = historical)) +
geom_line(aes(y = .value, group = paste(historical, .draw)),
alpha = .1) +
geom_point(data = equality, size = 0.1)
```

Unlike the prior plausible models in Figure 12.7 which were all over the place, the messages are clear.
At any urban population level, historically `dem`

states tend to have the most anti-discrimination laws and `gop`

states the fewest.
Further, the number of laws in a state tend to increase with urban percentage.
To dig into the details, we can examine the posterior models for the regression parameters \(\beta_0\) `(Intercept)`

, \(\beta_1\) (`percent_urban`

), \(\beta_2\) (`historicalgop`

), and \(\beta_3\) (`historicalswing`

):

```
tidy(equality_model, conf.int = TRUE, conf.level = 0.80)
# A tibble: 4 x 5
term estimate std.error conf.low conf.high<chr> <dbl> <dbl> <dbl> <dbl>
1 (Intercept) 1.71 0.303 1.31 2.09
2 percent_urban 0.0164 0.00353 0.0119 0.0210
3 historicalgop -1.52 0.134 -1.69 -1.34
4 historicalswing -0.610 0.103 -0.745 -0.477
```

Thus the posterior median relationship of a state’s number of anti-discrimination laws with its urban population and historical voting trends can be described by:

\[\begin{equation} \begin{split} \log(\lambda_i) & = 1.71 + 0.0164 X_{i1} - 1.52 X_{i2} - 0.61 X_{i3} \\ \lambda_i & = e^{1.71 + 0.0164 X_{i1} - 1.52 X_{i2} - 0.61 X_{i3}} \\ \end{split} \tag{12.4} \end{equation}\]

Consider the `percent_urban`

coefficient \(\beta_1\) which has a posterior median of roughly 0.0164.
Then when controlling for `historical`

voting trends, we expect the *logged* number of anti-discrimination laws in states to increase by 0.0164 for every extra percentage point in urban population.
More meaningfully (on the *unlogged* scale), if the urban population in one state is 1 percentage point greater than another state, we’d expect it to have 1.0165 *times* the number of, or 1.65% more, anti-discrimination laws:

\[e^{0.0164} = 1.0165 \;.\]

Or, if the urban population in one state is *25* percentage points greater than another state, we’d expect it to have roughly *one and a half* times the number of, or 51% more, anti-discrimination laws (\(e^{25 \cdot 0.0164} = 1.5068\)).
Take a quick quiz to similarly interpret the \(\beta_3\) coefficient for historically `swing`

states.^{60}

The posterior median of \(\beta_3\) is roughly -0.61. Correspondingly, \(e^{-0.61}\) \(=\) 0.54. How can we interpret this value? When holding constant `percent_urban`

…

- The number of anti-discrimination laws tends to decrease by roughly 54 percent for every extra
`swing`

state. `swing`

states tend to have 54 percent as many anti-discrimination laws as`dem`

leaning states.`swing`

states tend to have 0.54 fewer anti-discrimination laws than`dem`

leaning states.

The key here is remembering that the categorical `swing`

state indicator provides a *comparison* to the `dem`

state reference level.
Then when controlling for a state’s urban population, we’d expect the *logged* number of anti-discrimination laws to be 0.61 *lower* in a `swing`

state than in a `dem`

leaning state.
Equivalently, on the *unlogged* scale, `swing`

states tend to have 54 *percent* as many anti-discrimination laws as `dem`

leaning states (\(e^{-0.61} = 0.54\)).

In closing out our posterior interpretation, notice that the 80% posterior credible intervals for (\(\beta_1, \beta_2, \beta_3)\) in the above `tidy()`

summary provide evidence that each coefficient significantly differs from 0.
For example, there’s an 80% posterior chance that the `percent_urban`

coefficient \(\beta_1\) is between 0.0119 and 0.021.
Thus when controlling for a state’s `historical`

political leanings, there’s a *significant* positive association between the number of anti-discrimination laws in a state and its urban population.
Further, when controlling for a state’s `percent_urban`

makeup, the number of anti-discrimination laws in `gop`

leaning and `swing`

states tend to be significantly *below* that of `dem`

leaning states – the 80% credible intervals for \(\beta_2\) and \(\beta_3\) both fall below 0.
These conclusions are consistent with the posterior plausible models in Figure 12.9.

## 12.4 Posterior prediction

To explore how the general Poisson model plays out in individual states, consider the state of Minnesota, a historically `dem`

state with 73.3% of residents residing in urban areas and 4 anti-discrimination laws:

```
%>%
equality filter(state == "minnesota")
# A tibble: 1 x 6
state region gop_2016 laws historical percent_urban<fct> <fct> <dbl> <dbl> <fct> <dbl>
1 minne… midwe… 44.9 4 dem 73.3
```

Based on the state’s demographics and political leanings, we can approximate a posterior predictive model for its number of anti-discrimination laws.
Importantly, reflecting the *Poisson* data structure of our model, the 20,000 simulated posterior predictions are *counts*:

```
# Calculate posterior predictions
set.seed(84735)
<- posterior_predict(
mn_prediction newdata = data.frame(percent_urban = 73.3,
equality_model, historical = "dem"))
head(mn_prediction, 3)
1
1,] 20
[2,] 17
[3,] 21 [
```

The resulting posterior predictive model anticipates that Minnesota has between 10 and 30 anti-discrimination laws (roughly), a range that falls far above the actual number of laws in that state (4).
This discrepancy reveals that a state’s demographics and political leanings, thus our Poisson model, don’t tell the full story behind its anti-discrimination laws.
It also reveals something about Minnesota itself.
For a state with such a high urban population and historically Democratic voting trends, it has an unusually small number of anti-discrimination laws.
(Again, the *number* of laws in a state isn’t necessarily a proxy for the overall *quality* of laws.)

```
mcmc_hist(mn_prediction, binwidth = 1) +
geom_vline(xintercept = 4) +
xlab("Predicted number of laws in Minnesota")
```

Recall that we needn’t use the `posterior_predict()`

shortcut function to simulate the posterior predictive model for Minnesota.
Rather, we could directly predict the number of laws in that state from each of the 20,000 posterior plausible parameter sets \(\left(\beta_0^{(i)}, \beta_1^{(i)}, \beta_2^{(i)}, \beta_3^{(i)}\right)\).
To this end, from each parameter set we:

Calculate the logged average number of laws in states like Minnesota, with a 73.3% urban percentage and historically Democrat voting patterns:

\[\log(\lambda^{(i)}) = \beta_0^{(i)} + \beta_1^{(i)}\cdot 73.3 + \beta_2^{(i)}\cdot0 + \beta_3^{(i)} \cdot 0\]

Transform \(\log(\lambda^{(i)})\) to obtain the (

*unlogged*) average number of laws in states like Minnesota, \(\lambda^{(i)}\); andSimulate a Pois\(\left(\lambda^{(i)}\right)\) outcome for the number of laws in Minnesota, \(Y_{\text{new}}^{(i)}\), using

`rpois()`

.

The result matches the shortcut!

```
# Predict number of laws for each parameter set in the chain
set.seed(84735)
as.data.frame(equality_model) %>%
mutate(log_lambda = `(Intercept)` + percent_urban*73.3 +
*0 + historicalswing*0,
historicalgoplambda = exp(log_lambda),
y_new = rpois(20000, lambda = lambda)) %>%
ggplot(aes(x = y_new)) +
stat_count()
```

## 12.5 Model evaluation

To close out our analysis, let’s evaluate the quality of our Poisson regression model of anti-discrimination laws.
The first two of our model evaluation questions are easy to answer.
**How fair is our model?**
Though we don’t believe there to be bias in the data collection process, we certainly do warn against mistaking the *general state-level trends* revealed by our analysis as cause to make assumptions about *individuals* based on their voting behavior or where they live.
Further, as we noted up front, an analysis of the *number* of state laws shouldn’t be confused with an analysis of the *quality* of state laws.
**How wrong is our model?**
Our `pp_check()`

in Figure 12.8 demonstrated that our Poisson regression assumptions are reasonable.

Consider the third model evaluation question: **How accurate are our model predictions?**
As we did for the state of Minnesota, we can examine the posterior predictive models for each of the 49 states in our data set.
The plot below illustrates the posterior predictive credible intervals for the number of laws in each state alongside the actual observed data.
Overall, our model does better at anticipating the number of laws in `gop`

states than in `dem`

or `swing`

states – more of the observed `gop`

data points fall within the bounds of their credible intervals.
As exhibited by the narrower intervals, we also have more posterior certainty about the `gop`

states where the number of laws tends to be consistently small.

```
# Simulate posterior predictive models for each state
set.seed(84735)
<- posterior_predict(equality_model, newdata = equality)
poisson_predictions
# Plot the posterior predictive models for each state
ppc_intervals_grouped(equality$laws, yrep = poisson_predictions,
x = equality$percent_urban,
group = equality$historical,
prob = 0.5, prob_outer = 0.95,
facet_args = list(scales = "fixed"))
```

We can formalize our observations from Figure 12.12 with a `prediction_summary()`

.
Across the 49 states in our study, the observed numbers of anti-discrimination laws tend to fall only 3.4 laws, or 1.3 posterior standard deviations, from their posterior mean predictions.
Given the range of the number of state laws, from 1 to 38, a typical prediction error of 3.4 is pretty good!
Countering this positive with a negative, the observed number of laws for only roughly 78% of the states fall within their corresponding 95% posterior prediction interval.
This means that our posterior predictive models “missed” or didn’t anticipate the number of laws in 22%, or 11, of the 49 states.

```
prediction_summary(model = equality_model, data = equality)
mae mae_scaled within_50 within_951 3.407 1.304 0.3265 0.7755
```

For due diligence, we also calculate the *cross-validated* posterior predictive accuracy in applying this model to a “new” set of 50 states, or the *same* set of 50 states under the recognition that our current data is just a random snapshot in time.
In this case, the results are similar, suggesting that our model is *not* overfit to our sample data – we expect it to perform just as well at predicting “new” states.

```
# Cross-validation
set.seed(84735)
<- prediction_summary_cv(model = equality_model,
poisson_cv data = equality, k = 10)
$cv
poisson_cv
mae mae_scaled within_50 within_951 3.788 1.264 0.3 0.725
```

## 12.6 Negative Binomial regression for overdispersed counts

In 2017, *Cards Against Humanity Saves America* launched a series of monthly surveys in order to get the “Pulse of the nation” (2017).
We’ll use their September 2017 poll results to model the number of books somebody has read in the past year (\(Y\)) by two predictors: their age (\(X_1\)) and their response to the question of whether they’d rather be *wise but unhappy* or *happy but unwise*:

\[X_2 = \begin{cases} 1 & \text{wise but unhappy} \\ 0 & \text{otherwise (i.e. happy but unwise)} \\ \end{cases}\]

Because we really don’t have any prior understanding of this relationship, we’ll utilize weakly informative priors throughout our analysis.
Moving on, let’s load and process the `pulse_of_the_nation`

data from the **bayesrules** package.
In doing so, we’ll remove some outliers, focusing on people that read fewer than 100 books:

```
# Load data
data(pulse_of_the_nation)
<- pulse_of_the_nation %>%
pulse filter(books < 100)
```

Figure 12.13 reveals some basic patterns in readership.
First, though most people read fewer than 11 books per year, there is a *lot* of variability in reading patterns from person to person.
Further, though there appears to be a very weak relationship between book readership and one’s age, readership appears to be slightly higher among people that would prefer wisdom over happiness (makes sense to us!).

```
ggplot(pulse, aes(x = books)) +
geom_histogram(color = "white")
ggplot(pulse, aes(y = books, x = age)) +
geom_point()
ggplot(pulse, aes(y = books, x = wise_unwise)) +
geom_boxplot()
```

Given the skewed, count structure of our `books`

variable \(Y\), the Poisson regression tool is a reasonable *first approach* to modeling readership by a person’s age and their prioritization of wisdom versus happiness:

```
<- stan_glm(
books_poisson_sim ~ age + wise_unwise,
books data = pulse, family = poisson,
prior_intercept = normal(0, 2.5, autoscale = TRUE),
prior = normal(0, 2.5, autoscale = TRUE),
prior_aux = exponential(1, autoscale = TRUE),
chains = 4, iter = 5000*2, seed = 84735)
```

BUT the results are definitely not great.
Check out the `pp_check()`

in Figure 12.14.

```
pp_check(books_poisson_sim) +
xlab("books")
```

Counter to the *observed* book readership, which is right skewed and tends to be below 11 books per year, the Poisson posterior simulations of readership are symmetric around 11 books year.
Simply put, the Poisson regression model is wrong.
*Why*?
Well, recall that Poisson regression preserves the Poisson property of equal mean and variance.
That is, it assumes that among subjects of similar age and perspectives on wisdom versus happiness (\(X_1\) and \(X_2\)), the *typical* number of books read is roughly equivalent to the *variability* in books read.
Yet the `pp_check()`

highlights that, counter to this assumption, we actually observe *high variability* in book readership relative to a *low average* readership.
We can confirm this observation with some numerical summaries.
First, the discrepancy in the mean and variance in readership is true across *all* subjects in our survey.
On average, people read 10.9 books per year, but the variance in book readership was a whopping 198 \(\text{books}^2\):

```
# Mean and variability in readership across all subjects
%>%
pulse summarize(mean = mean(books), var = var(books))
# A tibble: 1 x 2
mean var<dbl> <dbl>
1 10.9 198.
```

When we `cut()`

the age range into three groups, we see that this is also true among subjects in the same general age bracket and with the same take on wisdom versus happiness, i.e. among subjects with similar \(X_1\) and \(X_2\) values.
For example, among respondents in the 45 to 72 year age bracket that prefer wisdom to happiness, the average readership was 12.5 books, a relatively small number in comparison to the variance of 270 \(\text{books}^2\).

```
# Mean and variability in readership
# among subjects of similar age and wise_unwise response
%>%
pulse group_by(cut(age,3), wise_unwise) %>%
summarize(mean = mean(books), var = var(books))
# A tibble: 6 x 4
# Groups: cut(age, 3) [3]
`cut(age, 3)` wise_unwise mean var
<fct> <fct> <dbl> <dbl>
1 (17.9,45] Happy but Unwise 9.23 138.
2 (17.9,45] Wise but Unhappy 12.6 195.
3 (45,72] Happy but Unwise 9.36 183.
4 (45,72] Wise but Unhappy 12.5 270.
5 (72,99.1] Happy but Unwise 12.6 236.
6 (72,99.1] Wise but Unhappy 10.2 97.0
```

In this case, we say that book readership \(Y\) is **overdispersed**.

**Overdispersion**

A random variable \(Y\) is **overdispersed** if the *observed* variability in \(Y\) exceeds the variability *expected* by the assumed probability model of \(Y\).

When our count response variable \(Y\) is too overdispersed to squeeze into the Poisson regression assumptions, we have some options.
Two common options, which produce similar results, are to (1) include an *overdispersion parameter* in the Poisson data model; or (2) utilize a non-Poisson data model.
Since it fits nicely into the modeling framework we’ve established, we’ll focus on the latter approach.
To this end, the **Negative Binomial probability model** is a useful alternative to the Poisson when \(Y\) is overdispersed.
*Like* the Poisson model, the Negative Binomial is suitable for count data \(Y \in \{0,1,2,\ldots\}\).
Yet *unlike* the Poisson, the Negative Binomial does *not* make the restrictive assumption that \(E(Y) = \text{Var}(Y)\).

**The Negative Binomial model**

Let random variable \(Y\) be some count, \(Y \in \{0,1,2,\ldots\}\), that can be modeled by the Negative Binomial with **mean parameter** \(\mu\) and **reciprocal dispersion parameter** \(r\):

\[Y | \mu, r \; \sim \; \text{NegBin}(\mu, r)\]

Then \(Y\) has conditional pmf

\[\begin{equation} f(y|\mu,r) = \left(\!\begin{array}{c} y + r - 1 \\ r \end{array}\!\right) \left(\frac{r}{\mu + r} \right)^r \left(\frac{\mu}{\mu + r} \right)^y\;\; \text{ for } y \in \{0,1,2,\ldots\} \tag{12.5} \end{equation}\]

with *unequal* mean and variance:

\[E(Y|\mu,r) = \mu \;\; \text{ and } \;\; \text{Var}(Y|\mu,r) = \mu + \frac{\mu^2}{r}\;.\]

**Comparisons to the Poisson model**

For *large* reciprocal dispersion parameters \(r\), \(\text{Var}(Y) \approx E(Y)\) and \(Y\) behaves much like a Poisson count variable.
For *small* \(r\), \(\text{Var}(Y) > E(Y)\) and \(Y\) is **overdispersed** in comparison to a Poisson count variable with the same mean.
Figure 12.15 illustrates these themes by example.

To make the switch to the more flexible Negative Binomial regression model of readership, we can simply swap out a Poisson data model for a Negative Binomial data model.
In doing so, we also pick up the extra reciprocal dispersion parameter, \(r > 0\), for which the Exponential provides a reasonable prior structure.
Our Negative Binomial regression model, along with the weakly informative priors scaled by `stan_glm()`

and obtained by `prior_summary()`

, follows:

\[\begin{equation} \begin{array}{lcrl} \text{data:} & \hspace{.025in} & Y_i|\beta_0,\beta_1,\beta_2,r & \stackrel{ind}{\sim} \text{NegBin}\left(\mu_i, r \right) \;\; \text{ with } \;\; \log\left( \mu_i \right) = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} \\ \text{priors:} & & \beta_{0c} & \sim N\left(0, 2.5^2\right) \\ & & \beta_1 & \sim N\left(0, 0.15^2\right) \\ & & \beta_2 & \sim N\left(0, 5.01^2\right) \\ & & r & \sim \text{Exp}(1)\\ \end{array} \tag{12.6} \end{equation}\]

Similarly, to simulate the posterior of regression parameters \((\beta_0,\beta_1,\beta_2,r)\), we can swap out `poisson`

for `neg_binomial_2`

in `stan_glm()`

:

```
<-
books_negbin_sim stan_glm(books ~ age + wise_unwise,
data = pulse, family = neg_binomial_2,
prior_intercept = normal(0, 2.5, autoscale = TRUE),
prior = normal(0, 2.5, autoscale = TRUE),
prior_aux = exponential(1, autoscale = TRUE),
chains = 4, iter = 5000*2, seed = 84735)
# Check out the priors
prior_summary(books_negbin_sim)
```

The results are fantastic. By incorporating more flexible assumptions about the variability in book readership, the posterior simulation of book readership very closely matches the observed behavior in our survey data (Figure 12.16). That is, the Negative Binomial regression model is not wrong (or, at least is much less wrong than the Poisson model).

```
pp_check(books_negbin_sim) +
xlim(0, 75) +
xlab("books")
```

With this peace of mind, we can continue just as we would with a Poisson analysis.
Mainly, since it utilizes a log transform, the interpretation of the Negative Binomial regression coefficients follows the same framework as in the Poisson setting.
Consider some posterior punchlines, supported by the `tidy()`

summaries below:

- When controlling for a person’s prioritization of wisdom versus happiness, there’s no significant association between age and book readership – 0 is squarely in the 80% posterior credible interval for
`age`

coefficient \(\beta_1\). - When controlling for a person’s age, people that prefer wisdom over happiness tend to read more than those that prefer happiness over wisdom – the 80% posterior credible interval for \(\beta_2\) is comfortably above 0. Assuming they’re the same age, we’d expect a person that prefers wisdom to read 1.3
*times*as many, or 30% more, books as somebody that prefers happiness (\(e^{0.266} = 1.3\)).

```
# Numerical summaries
tidy(books_negbin_sim, conf.int = TRUE, conf.level = 0.80)
# A tibble: 3 x 5
term estimate std.error conf.low conf.high<chr> <dbl> <dbl> <dbl> <dbl>
1 (Intercept) 2.23 0.131 2.07 2.41
2 age 0.000365 0.00239 -0.00270 0.00339
3 wise_unwiseWis… 0.266 0.0798 0.162 0.368
```

## 12.7 Generalized linear models: building on the theme

Though the Normal, Poisson, and Negative Binomial data structures are common among Bayesian regression models, we’re not limited to just these options.
We can also use `stan_glm()`

to fit models with Binomial, Gamma, inverse Normal, and other data structures.
*All* of these options belong to a larger class of **generalized linear models** (GLMs).

We needn’t march through every single GLM to consider them part of our toolbox.
We can build a GLM with *any* of the above data structures by drawing upon the principles we’ve developed throughout Unit 3.
First, it’s important to *note the structure in our response variable \(Y\)*.
Is \(Y\) discrete or continuous?
Symmetric or skewed?
What range of values can \(Y\) take?
These questions can help us identify an appropriate data structure.
Second, let \(E(Y|\ldots)\) denote the average \(Y\) value as defined by its data structure.
For all GLMs, the dependence of \(E(Y|\ldots)\) on a linear combination of predictors \((X_1, X_2, ..., X_p)\) is expressed by

\[g\left(E(Y|\ldots)\right) = \beta_0 + \beta_1 X_1 + \beta_2 X_2 + \cdots + \beta_p X_p\]

where the appropriate **link function** \(g(\cdot)\) depends upon the data structure.
For example, in Normal regression, the data is modeled by

\[Y_i | \beta_0, \beta_1, \cdots, \beta_p, \sigma \sim N(\mu_i, \sigma^2)\]

and the dependence of \(E(Y_i | \beta_0, \beta_1, \cdots, \beta_p, \sigma) = \mu_i\) on the predictors by

\[g(\mu_i) = \mu_i = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \cdots + \beta_p X_{ip} \; .\]

Thus Normal regression utilizes an **identity link function** since \(g(\mu_i)\) is equal to \(\mu_i\) itself.
In Poisson regression, the count data is modeled by

\[Y_i | \beta_0, \beta_1, \cdots, \beta_p \sim Pois(\lambda_i)\]

and the dependence of \(E(Y_i | \beta_0, \beta_1, \cdots, \beta_p) = \lambda_i\) on the predictors by

\[g(\lambda_i) := \log(\lambda_i) = \beta_0 + \beta_1 X_{i1} + \beta_2 X_{i2} + \cdots + \beta_p X_{ip} \; .\]

Thus Poisson regression utilizes a **log link function** since \(g(\lambda_i) = \log(\lambda_i)\).
The same is true for Negative Binomial regression.
We’ll dig into one more GLM, **logistic regression**, in the next chapter.
We hope that our survey of these four specific tools (Normal, Poisson, Negative Binomial, and logistic regression) empowers you to implement other GLMs in your own Bayesian practice.

## 12.8 Chapter summary

Let response variable \(Y \in \{0,1,2,\ldots\}\) be a discrete *count* of events that occur in a fixed interval of time or space.
In this context, using Normal regression to model \(Y\) by predictors \((X_1,X_2,\ldots,X_p)\) is often inappropriate – it assumes that \(Y\) is symmetric and can be negative.
The **Poisson regression model** offers a promising alternative:

\[\begin{equation} \begin{array}{lcrl} \text{data:} & \hspace{.025in} & Y_i|\beta_0,\beta_1,\ldots,\beta_p & \stackrel{ind}{\sim} Pois\left(\lambda_i \right) \;\; \text{ with } \;\; \log\left( \lambda_i \right) = \beta_0 + \beta_1 X_{i1} + \cdots + \beta_p X_{ip} \\ \text{priors:} & & \beta_{0c} & \sim N(m_0, s_0^2) \\ & & \beta_{1} & \sim N(m_1, s_1^2) \\ && \vdots & \\ & & \beta_{p} & \sim N(m_p, s_p^2) \\ \end{array} \tag{12.7} \end{equation}\]

One major constraint of Poisson regression is its assumption that, at any set of predictor values, the typical value of \(Y\) and variability in \(Y\) are equivalent.
Thus when \(Y\) is **overdispersed**, i.e. its variability exceeds assumptions, we might instead utilize the more flexible **Negative Binomial regression model**:

\[\begin{equation} \begin{array}{lcrl} \text{data:} & \hspace{.025in} & Y_i|\beta_0,\beta_1,\ldots,\beta_p,r & \stackrel{ind}{\sim} \text{NegBin}\left(\mu_i, r \right) \;\; \text{ with } \;\; \log\left( \mu_i \right) = \beta_0 + \beta_1 X_{i1} + \cdots + \beta_p X_{ip} \\ \text{priors:} & & \beta_{0c} & \sim N(m_0, s_0^2) \\ & & \beta_{1} & \sim N(m_1, s_1^2) \\ && \vdots & \\ & & \beta_{p} & \sim N(m_p, s_p^2) \\ & & r & \sim \text{Exp}(\ldots)\\ \end{array} \tag{12.8} \end{equation}\]

## 12.9 Exercises

### 12.9.1 Conceptual exercises

**Exercise 12.1 (Warming up)**

- Give a
*new*example (i.e. not the same as from the chapter) in which we would want to use a Poisson, instead of Normal, regression model. - The Poisson regression model uses a log link function, while the Normal regression model uses an identity link function. Explain in one or two sentences what a link function is.
- Explain why the log link function is used in Poisson regression.
- List the four assumptions for a Poisson regression model.

**Exercise 12.2 (Poisson versus Negative Binomial)**Specify whether Poisson regression, Negative Binomial regression, both, or neither fit with each situation described below.

- The response variable is a count.
- The link is a log.
- The link is the identity.
- We need to account for overdispersion.
- The response is a count variable, and as the expected response increases, the variability also increases.

**Exercise 12.3 (Why use a Negative Binomial)**You and your friend Nico are in a two-person

*Bayes Rules!*book club. How lovely! Nico has read only part of this chapter, and now they know about about Poisson regression, but not Negative Binomial regression. Be a good friend and answer their questions.

- What’s the shortcoming of Poisson regression?
- How does Negative Binomial regression address the shortcoming of Poisson regression?
- Are there any situations in which Poisson regression would be a better choice than Negative Binomial?

**Exercise 12.4 (Interpreting Poisson regression coefficients) **As modelers, the ability to interpret regression coefficients is of utmost importance. Let \(Y\) be the number of “Likes” a tweet gets in an hour, \(X_1\) be the number of followers the person who wrote the tweet has, and \(X_2\) indicate whether the tweet includes an emoji (\(X_2=1\) if there is an emoji, \(X_2= 0\) if there is no emoji). Further, suppose \(Y|\beta_0, \beta_1, \beta_2 \sim \text{Pois}(\lambda)\) with

- Interpret \(e^{\beta_0}\) in context.
- Interpret \(e^{\beta_1}\) in context.
- Interpret \(e^{\beta_2}\) in context.
- Provide the model equation for the expected number of “Likes” for a tweet in one hour, when the person who wrote the tweet has 300 followers, and the tweet does not use an emoji.

### 12.9.2 Applied exercises

**Exercise 12.5 (Eagles: get to know the data)**In the next exercises, you will explore how the number of eagle sightings in Ontario, Canada has changed over time. Since this context is unfamiliar to us, we’ll utilize weakly informative priors throughout. We’ll balance this prior uncertainty by the

`bald_eagles`

data in the **bayesrules**package which includes data on bald eagle sightings during 37 different one-week observation periods. First, get to know this data.

- Construct and discuss a univariate plot of
`count`

, the number of eagle sightings across the observation periods. - Construct and discuss a plot of
`count`

versus`year`

. - In exploring the number of eagle sightings over time, it’s important to consider the fact that the length of the observation periods vary from year to year, ranging from 134 to 248.75 hours. Update your plot from part b to also include information about the observation length in
`hours`

and comment on your findings.

**Exercise 12.6 (Eagles: First model attempt)**Our next goal is to model the relationship between bald eagle counts \(Y\) by year \(X_1\) when controlling for the number of observation hours \(X_2\). To begin, consider a

*Normal*regression model of \(Y\) versus \(X_1\) and \(X_2\).

- Simulate the model posterior and check the
`prior_summary()`

. - Use careful notation to write out the complete Bayesian structure of the Normal regression model of \(Y\) by \(X_1\) and \(X_2\).
- Complete a
`pp_check()`

for the Normal model. Use this to explain whether the model is “good” and, if not, what assumptions it makes that are inappropriate for the bald eagle analysis.

**Exercise 12.7 (Eagles: Second model attempt)**Let’s try to do better. Consider a

*Poisson*regression model of \(Y\) versus \(X_1\) and \(X_2\).

- In the bald eagle analysis, why might a
*Poisson*regression approach be more appropriate than a*Normal*regression approach? - Simulate the posterior of the Poisson regression model of \(Y\) versus \(X_1\) and \(X_2\). Check the
`prior_summary()`

. - Use careful notation to write out the complete Bayesian structure of the Poisson regression model of \(Y\) by \(X_1\) and \(X_2\).
- Complete a
`pp_check()`

for the Poisson model. Use this to explain whether the model is “good” and, if not, what assumptions it makes that are inappropriate for the bald eagle analysis.

**Exercise 12.8 (Eagles: An even better model)**The Poisson regression model of bald eagle counts (\(Y\)) by year (\(X_1\)) and observation hours (\(X_2\)), was pretty good. Let’s see if a Negative Binomial approach is even better.

- Simulate the model posterior and use a
`pp_check()`

to confirm whether the Negative Binomial model is reasonable. - Use careful notation to write out the complete Bayesian structure of the Negative Binomial regression model of \(Y\) by \(X_1\) and \(X_2\).
- Interpret the posterior median estimates of the regression coefficients on
`year`

and`hours`

, \(\beta_1\) and \(\beta_2\). Do so on the unlogged scale. - Construct and interpret a 95% posterior credible interval for the
`year`

coefficient. - When controlling for the number of observation hours, do we have ample evidence that the rate of eagle sightings has increased over time?

**Exercise 12.9 (Eagles: Model evaluation)**Finally, let’s evaluate the quality of our Negative Binomial bald eagle model.

- How fair is the model?
- How wrong is the model?
- How accurate are the model predictions?

**Exercise 12.10 (Open exercise: AirBnB)**The

`airbnb_small`

data in the **bayesrules**package contains information on AirBnB rentals in Chicago. This data was originally collated by Trinh and Ameri (2016) and distributed by Legler and Roback (2020). In this open-ended exercise, build, interpret, and evaluate a model of the number of

`reviews`

an AirBnB property has by its `rating`

, `district`

, `room_type`

, and the number of guests it `accommodates`

.
### References

*Beyond Multiple Linear Regression: Applied Generalized Linear Models and Multilevel Models in R*. CRC Press.

*Project for Statistics 316-Advanced Statistical Modeling, St. Olaf College*.

*Human Rights Campaign Foundation*.