# Chapter 09: Models for the Binomial family

    library(glmbayes)

## 1. Introductory Discussion

Binomial generalized linear models (GLMs) are used when the response represents binary outcomes (success/failure) or proportions (successes out of trials). They are among the most widely used GLMs in applied statistics, powering models for:

- clinical trial response rates
- yes/no survey outcomes
- conversion rates in marketing
- event occurrence indicators
- grouped binomial data (successes out of n)

Binomial regression is a standard generalized linear model (Nelder and Wedderburn 1972; McCullagh and Nelder 1989; Agresti 2015).

In classical statistics, these models are fit using:

    glm(..., family = binomial(link = ...))

In glmbayes, the Bayesian analogue is:

    glmb(..., family = binomial(link = ...), pfamily = dNormal(mu, Sigma))

This chapter introduces:

1. the structure of binomial GLMs
2. the available link functions (logit, probit, cloglog)
3. how to specify these models in glmbayes
4. worked examples for each link function

We build on the foundations from Chapters 07 and 08, especially the role of link functions, log‑concavity, and prior specification.

## 2. Binomial Likelihood and Weighted Formulation

Binomial data arise in several equivalent representations:

- Binary outcomes: \(y_i \in \{0,1\}\)
- Proportions: \(y_i = \text{successes}_i / n_i\) with weights \(w_i = n_i\)
- Two‑column matrix: \(\text{cbind(successes, failures)}\)

In all cases, the underlying sampling model is

\[ Y_i \sim \text{Binomial}(n_i, \mu_i), \qquad 0 < \mu_i < 1, \]

where: - \(n_i\) is the number of trials, - \(\mu_i = \Pr(Y_i = 1)\) is the success probability.

### 2.1 Linear predictor and mean structure

A binomial GLM links the mean \(\mu_i\) to a linear predictor through

\[ \eta_i = x_i^\top \beta, \qquad \mu_i = g^{-1}(\eta_i), \]

where \(g(\cdot)\) is the chosen link function (logit, probit, cloglog, etc.).

### 2.2 Weighted binomial log‑likelihood

Using weights \(w_i = n_i\), the log‑likelihood (up to constants) becomes

\[ \ell(\beta) = \sum_{i=1}^n w_i\Big[ y_i \log(\mu_i) + (1-y_i)\log(1-\mu_i) \Big]. \]

This form is used by both `glm()` and the Bayesian functions `glmb()` and `rglmb()`.

### 2.3 Exponential‑family representation

The binomial likelihood belongs to the exponential family (McCullagh and Nelder 1989; Agresti 2015).
For a model with linear predictor \[ \eta_i = x_i^\top \beta, \] and mean \[ \mu_i = g^{-1}(\eta_i), \] the contribution of observation \(i\) to the log‑likelihood can be written as \[ \ell_i(\beta) = w_i\Big[ y_i \log(\mu_i) + (1-y_i)\log(1-\mu_i) \Big], \] where \(w_i\) is the number of trials (or a user‑supplied weight).

This representation does not require the link to be canonical.
The variance of a binomial observation is always \[ \mathrm{Var}(Y_i) = \mu_i(1-\mu_i), \] regardless of the link function.

### 2.4 Link functions

The link function determines how the mean response relates to the linear predictor: \[ g(\mu_i) = \eta_i. \]

The `binomial()` family in R supports several link functions:

Link Function | Formula | Notes
--- | --- | ---
logit (canonical) | \(\eta = \log(\mu/(1-\mu))\) | Most common; canonical link; induces log‑concavity in \(\eta\)
probit | \(\eta = \Phi^{-1}(\mu)\) | Based on the standard normal CDF; induces log‑concavity in \(\eta\)
cloglog | \(\eta = \log[-\log(1-\mu)]\) | Asymmetric; useful for rare events; induces log‑concavity in \(\eta\)
cauchit | \(\eta = \tan[\pi(\mu - 1/2)]\) | Heavy‑tailed; does not preserve log‑concavity in general
identity | \(\eta = \mu\) | Must ensure \(0 < \mu < 1\); does not preserve log‑concavity in general

In this chapter we focus on the three most commonly used links:

- logit (Section 4)
- probit (Section 5)
- cloglog (Section 6)

Each link produces a different transformation \(g^{-1}(\eta)\) and therefore a different expression for the log‑likelihood and its derivatives. For the logit, probit, and cloglog links, the resulting log‑likelihood is known to be log‑concave in the linear predictor \(\eta\), which is crucial for stable envelope construction and accept–reject sampling in glmbayes.

The explicit formulas for each link are provided at the beginning of their respective sections.

## 3. Specifying Binomial Models in glmbayes

The general Bayesian call is:

    glmb(
      formula,
      family   = binomial(link = "logit" | "probit" | "cloglog"),
      pfamily  = dNormal(mu = mu, Sigma = V),
      data     = ...
    )

### 3.1 Prior Specification

As in earlier chapters, the recommended workflow is:

    ps <- Prior_Setup(formula, family = binomial(link = "logit"), data = ...)
    mu <- ps$mu
    V  <- ps$Sigma

This produces:

- a Zellner‑type g‑prior (Griffin and Brown 2010)
- prior means aligned with the null model
- prior variances scaled relative to the likelihood curvature

You may override these defaults for more informative priors (see Chapter 14).

## 4. Logit Link: Likelihood, Prior, Posterior

For the logit link: \[ \eta_i = \log\!\left(\frac{\mu_i}{1-\mu_i}\right), \qquad \mu_i = \frac{1}{1+e^{-\eta_i}}. \]

### Weighted Log-Likelihood

\[ \ell_{\text{logit}}(\beta) = \sum_{i=1}^n w_i\Big[ y_i\,\eta_i - \log(1+e^{\eta_i}) \Big]. \]

### Normal Prior

\[ \log p(\beta) = -\tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

### Log-Posterior

\[ \log p(\beta \mid y) = \sum_{i=1}^n w_i\Big[ y_i\,\eta_i - \log(1+e^{\eta_i}) \Big] - \tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

The logit link is the canonical choice for binomial GLMs (McCullagh and Nelder 1989; Agresti 2015):

\[ \eta = \log\left(\frac{\mu}{1-\mu}\right) \]

It is symmetric and interpretable in terms of log‑odds.

### 4.1 Example Data

We use the Menarche dataset introduced in Chapter 08:

    data(menarche,package="MASS")
    Age2 <- menarche$Age - 13
    Menarche_Model_Data <- data.frame(
      Age = menarche$Age,
      Total = menarche$Total,
      Menarche = menarche$Menarche,
      Age2 = Age2
    )
    Menarche_Model_Data
    #>      Age Total Menarche  Age2
    #> 1   9.21   376        0 -3.79
    #> 2  10.21   200        0 -2.79
    #> 3  10.58    93        0 -2.42
    #> 4  10.83   120        2 -2.17
    #> 5  11.08    90        2 -1.92
    #> 6  11.33    88        5 -1.67
    #> 7  11.58   105       10 -1.42
    #> 8  11.83   111       17 -1.17
    #> 9  12.08   100       16 -0.92
    #> 10 12.33    93       29 -0.67
    #> 11 12.58   100       39 -0.42
    #> 12 12.83   108       51 -0.17
    #> 13 13.08    99       47  0.08
    #> 14 13.33   106       67  0.33
    #> 15 13.58   105       81  0.58
    #> 16 13.83   117       88  0.83
    #> 17 14.08    98       79  1.08
    #> 18 14.33    97       90  1.33
    #> 19 14.58   120      113  1.58
    #> 20 14.83   102       95  1.83
    #> 21 15.08   122      117  2.08
    #> 22 15.33   111      107  2.33
    #> 23 15.58    94       92  2.58
    #> 24 15.83   114      112  2.83
    #> 25 17.58  1049     1049  4.58

### 4.2 Prior Setup

    ps <- Prior_Setup(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family = binomial(link = "logit"),
      data = Menarche_Model_Data
    )
    #> Using default pwt = 0.01 (low-d default).
    mu <- ps$mu
    V  <- ps$Sigma

### 4.3 Fit the Model

    glmb.logit <- glmb(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family  = binomial(link = "logit"),
      pfamily = dNormal(mu = mu, Sigma = V),
      data    = Menarche_Model_Data,
      n       = 1000
    )

### 4.4 Summary

    summary(glmb.logit)
    #> Call
    #> glmb(formula = cbind(Menarche, Total - Menarche) ~ Age2, family = binomial(link = "logit"), pfamily = dNormal(mu = mu, Sigma = V), n = 1000, data = Menarche_Model_Data)
    #>
    #> Expected Residuals:
    #>        Min         1Q     Median         3Q        Max
    #> -2.0755848 -1.0387660 -0.4569500  0.7209114  1.3238875
    #>
    #> Prior and Maximum Likelihood Estimates with Standard Deviations
    #>
    #>             Null Mode Prior Mean Prior.sd Max Like. Like.sd
    #> (Intercept)   0.36015    0.36015  0.62794  -0.01081   0.063
    #> Age2          0.00000    0.00000  0.58658   1.63197   0.059
    #>
    #> Bayesian Estimates Based on 1000 iid draws
    #>
    #>             Post.Mode Post.Mean   Post.Sd  MC Error Pr(Null_tail) SE(tail)
    #> (Intercept) -0.007133 -0.007677  0.059553  0.001883      0.000000        0
    #> Age2         1.615870  1.622010  0.058152  0.001839      0.000000        0
    #>             Pr(Prior_tail)
    #> (Intercept)         <2e-16 ***
    #> Age2                <2e-16 ***
    #> ---
    #> Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
    #>   Directional Tail Summaries:
    #>
    #>                Metric vs Null vs Prior
    #>  Mahalanobis Distance 28.5021  28.5021
    #>      Tail Probability  0.0000   0.0000
    #>   [Tail probabilities are P(delta^T * Z <= 0) in whitened space]
    #>
    #>
    #> Distribution Percentiles
    #>
    #>                  1.0%      2.5%      5.0%    Median     95.0%     97.5% 99.0%
    #> (Intercept) -0.138670 -0.124034 -0.107398 -0.007492  0.090212  0.111000 0.131
    #> Age2         1.494805  1.510505  1.524865  1.621645  1.718166  1.731505 1.750
    #>
    #> Effective Number of Parameters: 1.882208
    #> Expected Residual Deviance: 28.6164
    #> DIC: 114.5504
    #>
    #> Expected Mean dispersion: 1
    #> Sq.root of Expected Mean dispersion: 1
    #>
    #> Mean Likelihood Subgradient Candidates Per iid sample: 1.286

This produces:

- posterior means and modes
- posterior standard deviations
- tail probabilities
- credible intervals
- DIC and effective parameter count (Spiegelhalter et al. 2002)

The slope parameter typically shows strong evidence of increasing probability of menarche with age.

## 5. Probit Link: Likelihood, Prior, Posterior

The probit link is a common alternative to the logit when a latent normal formulation is convenient (McCullagh and Nelder 1989; Agresti 2015).

For the probit link: \[ \eta_i = \Phi^{-1}(\mu_i), \qquad \mu_i = \Phi(\eta_i), \] where \(\Phi\) is the standard normal CDF.

### Weighted Log-Likelihood

\[ \ell_{\text{probit}}(\beta) = \sum_{i=1}^n w_i\Big[ y_i \log\Phi(\eta_i) + (1-y_i)\log\big(1-\Phi(\eta_i)\big) \Big]. \]

### Normal Prior

\[ \log p(\beta) = -\tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

### Log-Posterior

\[ \log p(\beta \mid y) = \sum_{i=1}^n w_i\Big[ y_i \log\Phi(\eta_i) + (1-y_i)\log\big(1-\Phi(\eta_i)\big) \Big] - \tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

It is similar to the logit link but has slightly thinner tails and a more Gaussian interpretation.

### 5.1 Prior Setup

    ps2 <- Prior_Setup(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family = binomial(link = "probit"),
      data = Menarche_Model_Data
    )
    #> Using default pwt = 0.01 (low-d default).
    mu2 <- ps2$mu
    V2  <- ps2$Sigma

### 5.2 Fit the Model

    glmb.probit <- glmb(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family  = binomial(link = "probit"),
      pfamily = dNormal(mu = mu2, Sigma = V2),
      data    = Menarche_Model_Data,
      n       = 1000
    )

### 5.3 Summary

    summary(glmb.probit)
    #> Call
    #> glmb(formula = cbind(Menarche, Total - Menarche) ~ Age2, family = binomial(link = "probit"), pfamily = dNormal(mu = mu2, Sigma = V2), n = 1000, data = Menarche_Model_Data)
    #>
    #> Expected Residuals:
    #>        Min         1Q     Median         3Q        Max
    #> -1.6345951 -0.9450241 -0.4905554  0.4574010  1.7831901
    #>
    #> Prior and Maximum Likelihood Estimates with Standard Deviations
    #>
    #>             Null Mode Prior Mean Prior.sd Max Like. Like.sd
    #> (Intercept)   0.22517    0.22517  0.34838  -0.01724   0.035
    #> Age2          0.00000    0.00000  0.29405   0.90782   0.030
    #>
    #> Bayesian Estimates Based on 1000 iid draws
    #>
    #>              Post.Mode  Post.Mean    Post.Sd   MC Error Pr(Null_tail) SE(tail)
    #> (Intercept) -0.0146442 -0.0162405  0.0346883  0.0010969     0.0000000        0
    #> Age2         0.8988270  0.8995796  0.0298859  0.0009451     0.0000000        0
    #>             Pr(Prior_tail)
    #> (Intercept)         <2e-16 ***
    #> Age2                <2e-16 ***
    #> ---
    #> Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
    #>   Directional Tail Summaries:
    #>
    #>                Metric vs Null vs Prior
    #>  Mahalanobis Distance 32.9300  32.9300
    #>      Tail Probability  0.0000   0.0000
    #>   [Tail probabilities are P(delta^T * Z <= 0) in whitened space]
    #>
    #>
    #> Distribution Percentiles
    #>
    #>                 1.0%     2.5%     5.0%   Median    95.0%    97.5% 99.0%
    #> (Intercept) -0.10246 -0.08880 -0.07313 -0.01517  0.03981  0.05282 0.065
    #> Age2         0.83459  0.84459  0.85097  0.89799  0.95012  0.95812 0.969
    #>
    #> Effective Number of Parameters: 2.027226
    #> Expected Residual Deviance: 24.99329
    #> DIC: 111.0723
    #>
    #> Expected Mean dispersion: 1
    #> Sq.root of Expected Mean dispersion: 1
    #>
    #> Mean Likelihood Subgradient Candidates Per iid sample: 1.289

The probit model typically yields:

- slightly smaller slope estimates
- slightly smaller posterior standard deviations
- similar fitted probabilities

The DIC often favors probit slightly for smooth S‑shaped curves (as seen in Chapter 05).

## 6. Complementary Log-Log (cloglog) Link: Likelihood, Prior, Posterior

The complementary log–log link is often used for asymmetric response curves and hazard‑type interpretations (McCullagh and Nelder 1989; Agresti 2015).

For the cloglog link: \[ \eta_i = \log\!\big[-\log(1-\mu_i)\big], \qquad \mu_i = 1 - \exp\!\big(-e^{\eta_i}\big). \]

### Weighted Log-Likelihood

\[ \ell_{\text{cloglog}}(\beta) = \sum_{i=1}^n w_i\Big[ y_i \log\!\big(1 - e^{-e^{\eta_i}}\big) + (1-y_i)\big(-e^{\eta_i}\big) \Big]. \]

### Normal Prior

\[ \log p(\beta) = -\tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

### Log-Posterior

\[ \log p(\beta \mid y) = \sum_{i=1}^n w_i\Big[ y_i \log\!\big(1 - e^{-e^{\eta_i}}\big) + (1-y_i)\big(-e^{\eta_i}\big) \Big] - \tfrac12(\beta-\mu_0)^\top \Sigma_0^{-1}(\beta-\mu_0) + \text{const}. \]

The cloglog link is asymmetric:

It is useful when:

- the event probability increases rapidly near 0
- the hazard interpretation is meaningful
- the response curve is skewed

### 6.1 Prior Setup

    ps3 <- Prior_Setup(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family = binomial(link = "cloglog"),
      data = Menarche_Model_Data
    )
    #> Using default pwt = 0.01 (low-d default).
    #> Warning: glm.fit: fitted probabilities numerically 0 or 1 occurred
    mu3 <- ps3$mu
    V3  <- ps3$Sigma

### 6.2 Fit the Model

    glmb.cloglog <- glmb(
      cbind(Menarche, Total - Menarche) ~ Age2,
      family  = binomial(link = "cloglog"),
      pfamily = dNormal(mu = mu3, Sigma = V3),
      data    = Menarche_Model_Data,
      n       = 1000
    )
    #> Warning: glm.fit: fitted probabilities numerically 0 or 1 occurred
    #> Warning: glm.fit: fitted probabilities numerically 0 or 1 occurred

### 6.3 Summary

    summary(glmb.cloglog)
    #> Call
    #> glmb(formula = cbind(Menarche, Total - Menarche) ~ Age2, family = binomial(link = "cloglog"), pfamily = dNormal(mu = mu3, Sigma = V3), n = 1000, data = Menarche_Model_Data)
    #>
    #> Expected Residuals:
    #>       Min        1Q    Median        3Q       Max
    #> -3.975469 -2.540734 -1.127176  1.192112  3.397299
    #>
    #> Prior and Maximum Likelihood Estimates with Standard Deviations
    #>
    #>             Null Mode Prior Mean Prior.sd Max Like. Like.sd
    #> (Intercept)   -0.1173    -0.1173   0.4094   -0.5960   0.041
    #> Age2           0.0000     0.0000   0.3117    0.9530   0.031
    #>
    #> Bayesian Estimates Based on 1000 iid draws
    #>
    #>              Post.Mode  Post.Mean    Post.Sd   MC Error Pr(Null_tail) SE(tail)
    #> (Intercept) -0.5910908 -0.5929338  0.0424279  0.0013417     0.0000000        0
    #> Age2         0.9451221  0.9466405  0.0286987  0.0009075     0.0000000        0
    #>             Pr(Prior_tail)
    #> (Intercept)         <2e-16 ***
    #> Age2                <2e-16 ***
    #> ---
    #> Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
    #>   Directional Tail Summaries:
    #>
    #>                Metric vs Null vs Prior
    #>  Mahalanobis Distance 33.1564  33.1564
    #>      Tail Probability  0.0000   0.0000
    #>   [Tail probabilities are P(delta^T * Z <= 0) in whitened space]
    #>
    #>
    #> Distribution Percentiles
    #>
    #>                1.0%    2.5%    5.0%  Median   95.0%   97.5%  99.0%
    #> (Intercept) -0.6992 -0.6784 -0.6620 -0.5936 -0.5233 -0.5099 -0.498
    #> Age2         0.8826  0.8897  0.8982  0.9470  0.9944  1.0029  1.011
    #>
    #> Effective Number of Parameters: 2.02536
    #> Expected Residual Deviance: 120.897
    #> DIC: 206.9741
    #>
    #> Expected Mean dispersion: 1
    #> Sq.root of Expected Mean dispersion: 1
    #>
    #> Mean Likelihood Subgradient Candidates Per iid sample: 1.291

The cloglog model often fits poorly for symmetric S‑shaped curves (as shown in Chapter 05), but it is valuable for:

- rare events
- survival‑type interpretations
- asymmetric growth curves

## 7. Comparing Link Functions

The Deviance Information Criterion (DIC) provides a Bayesian analogue to AIC (Spiegelhalter et al. 2002):

    DIC_comp<-rbind(
      extractAIC(glmb.logit),
      extractAIC(glmb.probit),
      extractAIC(glmb.cloglog))

    rownames(DIC_comp)<-c("logit","probit","cloglog")
    DIC_comp
    #>               pD      DIC
    #> logit   1.882208 114.5504
    #> probit  2.027226 111.0723
    #> cloglog 2.025360 206.9741

Typical patterns:

- logit and probit often perform similarly
- probit may slightly outperform logit for smooth curves
- cloglog is best when the response is asymmetric or hazard‑like

The effective number of parameters (pD) is part of the same framework (Spiegelhalter et al. 2002) and helps diagnose model complexity.

## 8. Concluding Discussion

Binomial GLMs are a core component of the glmbayes package. Their log‑concave likelihoods make them ideal for the envelope‑based accept‑reject sampler, and the familiar link functions allow analysts to choose models that match the scientific context (McCullagh and Nelder 1989; Gelman et al. 2013).

This chapter demonstrated:

- how binomial GLMs are structured
- how link functions differ
- how to specify priors using Prior_Setup
- how to fit logit, probit, and cloglog models
- how to compare models using DIC

In the next chapter, we extend these ideas to Poisson models, which share many structural similarities but introduce new considerations for count data.

# Appendix A. Bayes Rules! companion — Perth rain (`weather_perth`)

Book: (Johnson et al. 2022), Chapter 13 — same model and informative priors as Chapter 08, Appendix A.
Model: `raintomorrow ~ humidity9am` (logit link)

> Requires `bayesrules`.

    library(bayesrules)
    weather <- bayesrules::weather_perth
    weather$raintomorrow <- as.integer(weather$raintomorrow == "Yes")
    mu_w <- matrix(c(-1.4, 0.07), nrow = 1)
    colnames(mu_w) <- c("(Intercept)", "humidity9am")
    Sigma_w <- diag(c(0.7^2, 0.035^2))
    dimnames(Sigma_w) <- list(colnames(mu_w), colnames(mu_w))
    book_br09 <- data.frame(
      parameter = c("(Intercept)", "humidity9am"),
      book_lo   = c(-5.08785, 0.04147),
      book_hi   = c(-4.13450, 0.05487),
      book_mid  = c(-4.611175, 0.04817),
      check.names = FALSE
    )

    set.seed(2026)
    glmb_rain <- glmb(
      raintomorrow ~ humidity9am,
      family  = binomial(),
      pfamily = dNormal(mu = mu_w, Sigma = Sigma_w),
      data    = weather,
      n       = 2000
    )
    #> [glmb_Standardize_Model][NOTE] Posterior Hessian is moderately ill-conditioned.
    #>   kappa(H) = 67316.9
    print(glmb_rain)
    #>
    #> Call:  glmb(formula = raintomorrow ~ humidity9am, family = binomial(),
    #>     pfamily = dNormal(mu = mu_w, Sigma = Sigma_w), n = 2000,
    #>     data = weather)
    #>
    #> Posterior Mean Coefficients:
    #> (Intercept)  humidity9am
    #>    -3.94628      0.03891
    #>
    #> Effective Number of Parameters: 1.76987
    #> Expected Residual Deviance: 874.5654
    #> DIC: 876.3353

    br09_compare <- data.frame(
      parameter = book_br09$parameter,
      `Book 80% lo` = book_br09$book_lo,
      `Book 80% hi` = book_br09$book_hi,
      `Book midpoint` = book_br09$book_mid,
      `glmb Post.Mean` = as.numeric(glmb_rain$coef.means[book_br09$parameter]),
      `glmb Post.Sd`   = sapply(book_br09$parameter, function(p)
        sd(glmb_rain$coefficients[, p, drop = TRUE])),
      check.names = FALSE
    )
    knitr::kable(br09_compare, digits = 4,
      caption = "Bayes Rules! Ch. 13 vs. glmb() (informative priors)")

Bayes Rules! Ch. 13 vs. glmb() (informative priors)
| parameter | Book 80% lo | Book 80% hi | Book midpoint | glmb Post.Mean | glmb Post.Sd
--- | --- | --- | --- | --- | --- | ---
(Intercept) | (Intercept) | -5.0879 | -4.1345 | -4.6112 | -3.9463 | 0.3172
humidity9am | humidity9am | 0.0415 | 0.0549 | 0.0482 | 0.0389 | 0.0046

Multivariable extensions (`humidity3pm`, `raintoday`) use `Prior_Setup()` defaults on extra coefficients unless you add further informative book priors (Chapter 14).

# References

Agresti, Alan. 2015. Foundations of Linear and Generalized Linear Models. Cambridge University Press.
Gelman, Andrew, John B. Carlin, Hal S. Stern, David B. Dunson, Aki Vehtari, and Donald B. Rubin. 2013. Bayesian Data Analysis. 3rd ed. CRC Press.
Griffin, Jim E., and Philip J. Brown. 2010. “Inference with Normal-Gamma Prior Distributions in Regression Problems.” Bayesian Analysis 5 (1): 171–88. https://doi.org/10.1214/10-BA507.
Johnson, Alicia A., Miles Q. Ott, and Mine Dogucu. 2022. Bayes Rules! An Introduction to Applied Bayesian Modeling. CRC Press. https://www.bayesrulesbook.com.
McCullagh, P., and J. A. Nelder. 1989. Generalized Linear Models. Chapman; Hall.
Nelder, J. A., and R. W. M. Wedderburn. 1972. “Generalized Linear Models.” Journal of the Royal Statistical Society. Series A (General) 135 (3): 370–84. https://doi.org/10.2307/2344614.
Spiegelhalter, David J., Nicky G. Best, Bradley P. Carlin, and Angelika van der Linde. 2002. “Bayesian Measures of Model Complexity and Fit.” Journal of the Royal Statistical Society: Series B (Statistical Methodology) 64 (4): 583–639. https://doi.org/10.1111/1467-9868.00353.