# 11 Fitting Basic Discrete-Time Hazard Models

In this chapter and the next, we present statistical models of hazard for data collected in discrete time. The relative simplicity of these models makes them an ideal entrée into the world of survival analysis. In subsequent chapters, we extend these basic ideas to situations in which event occurrence is recorded in continuous time.

Good data analysis involves more than using a computer package to fit a statistical model to data. To conduce a credible discrete-time survival analysis, you must: (1) specify a suitable model for hazard and understand its assumptions; (2) use sample data to estimate the model parameters; (3) interpret results in terms of your research questions; (4) evaluate model fit and [express the uncertainty in the] model parameters; and (5) communicate your findings. (Singer & Willett, 2003, pp. 357–358)

## 11.1 Toward a statistical model for discrete-time hazard

Time to load Capaldi, Crosby, and Stoolmiller’s (1996) `firstsex.csv`

data.

```
library(tidyverse)
<- read_csv("data/firstsex.csv")
sex
glimpse(sex)
```

```
## Rows: 180
## Columns: 5
## $ id <dbl> 1, 2, 3, 5, 6, 7, 9, 10, 11, 12, 13, 14, 15, 16, 17, 18, 19, 20, 21, 22, 23, 24, 25, 26, 28, …
## $ time <dbl> 9, 12, 12, 12, 11, 9, 12, 11, 12, 11, 12, 11, 9, 12, 10, 12, 7, 12, 10, 12, 11, 12, 11, 12, 1…
## $ censor <dbl> 0, 1, 1, 0, 0, 0, 1, 0, 1, 0, 1, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 1, 0, 1, 0, 1, 0, 0, 0, 0, 0, …
## $ pt <dbl> 0, 1, 0, 1, 0, 1, 0, 0, 1, 1, 0, 1, 1, 1, 1, 1, 0, 0, 1, 1, 1, 0, 0, 0, 1, 1, 1, 0, 1, 1, 0, …
## $ pas <dbl> 1.9788670, -0.5454916, -1.4049800, 0.9741806, -0.6356313, -0.2428857, -0.8685001, 0.4535947, …
```

Here are the cases broken down by `time`

and `censor`

status.

```
%>%
sex count(time, censor)
```

```
## # A tibble: 7 × 3
## time censor n
## <dbl> <dbl> <int>
## 1 7 0 15
## 2 8 0 7
## 3 9 0 24
## 4 10 0 29
## 5 11 0 25
## 6 12 0 26
## 7 12 1 54
```

Since these data show no censoring before the final `time`

point, it is straightforward to follow along with the text (p. 358) and compute the percent who had already had sex by the 12^{th} grade.

```
%>%
sex count(censor) %>%
mutate(percent = 100 * (n / sum(n)))
```

```
## # A tibble: 2 × 3
## censor n percent
## <dbl> <int> <dbl>
## 1 0 126 70
## 2 1 54 30
```

Here we break the data down by our central predictor, `pt`

, which is coded “0 for boys who lived with both biological parents” and “1 for boys who experienced one or more parenting transitions” before the 7^{th} grade.

```
%>%
sex count(pt) %>%
mutate(percent = 100 * (n / sum(n)))
```

```
## # A tibble: 2 × 3
## pt n percent
## <dbl> <int> <dbl>
## 1 0 72 40
## 2 1 108 60
```

### 11.1.1 Plots of within-group hazard functions and survivor functions.

Plots of sample hazard functions and survivor functions estimates separately for groups distinguished by their predictor values are invaluable exploratory tools. If a predictor is categorical, like

PT, construction of these displays is straightforward. If a predictor is continuous, you should just temporarily categorize its values for plotting purposes. (pp. 358–359,emphasisin the original)

To make our version of the descriptive plots in Figure 11.1, we’ll need to first load the **survival** package.

`library(survival)`

`fit11.1`

will be of the cases for which `pt == 0`

and `fit11.2`

will be of the cases for which `pt == 1`

. With `fit11.3`

, we use all cases regardless of `pt`

status.

```
.1 <-
fit11survfit(data = sex %>% filter(pt == 0),
Surv(time, 1 - censor) ~ 1)
.2 <-
fit11survfit(data = sex %>% filter(pt == 1),
Surv(time, 1 - censor) ~ 1)
.3 <-
fit11survfit(data = sex,
Surv(time, 1 - censor) ~ 1)
```

Before we plot the results, it might be handy to arrange the fit results in life tables. We can streamline that code with the custom `make_lt()`

function from last chapter.

```
<- function(fit) {
make_lt
# arrange the lt data for all rows but the first
<-
most_rows tibble(time = fit$time) %>%
mutate(time_int = str_c("[", time, ", ", time + 1, ")"),
n_risk = fit$n.risk,
n_event = fit$n.event) %>%
mutate(n_censored = n_risk - n_event - lead(n_risk, default = 0),
hazard_fun = n_event / n_risk,
survivor_fun = fit$surv)
# define the values for t = 2 and t = 1
<- fit$time[1]
time_1 <- time_1 - 1
time_0
# define the values for the row for which t = 1
<-
row_1 tibble(time = time_0,
time_int = str_c("[", time_0, ", ", time_1, ")"),
n_risk = fit$n.risk[1],
n_event = NA,
n_censored = NA,
hazard_fun = NA,
survivor_fun = 1)
# make the full life table
<-
lt bind_rows(row_1,
most_rows)
lt
}
```

We’ll use `make_lt()`

separately for each fit, stack the results from the first on top of those from the second, and add a `pt`

column to index the rows. This will be our version of Table 11.1 (p. 360).

```
<-
lt bind_rows(make_lt(fit11.1),
make_lt(fit11.2),
make_lt(fit11.3)) %>%
mutate(pt = factor(rep(c("pt = 0", "pt = 1", "overall"), each = n() / 3))) %>%
select(pt, everything())
lt
```

```
## # A tibble: 21 × 8
## pt time time_int n_risk n_event n_censored hazard_fun survivor_fun
## <fct> <dbl> <chr> <dbl> <dbl> <dbl> <dbl> <dbl>
## 1 pt = 0 6 [6, 7) 72 NA NA NA 1
## 2 pt = 0 7 [7, 8) 72 2 0 0.0278 0.972
## 3 pt = 0 8 [8, 9) 70 2 0 0.0286 0.944
## 4 pt = 0 9 [9, 10) 68 8 0 0.118 0.833
## 5 pt = 0 10 [10, 11) 60 8 0 0.133 0.722
## 6 pt = 0 11 [11, 12) 52 10 0 0.192 0.583
## 7 pt = 0 12 [12, 13) 42 8 34 0.190 0.472
## 8 pt = 1 6 [6, 7) 108 NA NA NA 1
## 9 pt = 1 7 [7, 8) 108 13 0 0.120 0.880
## 10 pt = 1 8 [8, 9) 95 5 0 0.0526 0.833
## # ℹ 11 more rows
```

Here is the code for the top panel of Figure 11.1.

```
<-
p1 %>%
lt filter(pt != "overall") %>%
ggplot(aes(x = time, y = hazard_fun, color = pt, group = pt)) +
geom_line() +
scale_color_viridis_d(NULL, option = "A", end = .5) +
scale_x_continuous("grade", breaks = 6:12, limits = c(6, 12)) +
scale_y_continuous("estimated hazard probability",
limits = c(0, .5)) +
theme(panel.grid = element_blank())
```

Now make the plot for the bottom panel.

```
<-
p2 %>%
lt filter(pt != "overall") %>%
ggplot(aes(x = time, y = survivor_fun, color = pt, group = pt)) +
geom_hline(yintercept = .5, color = "white", linetype = 2) +
geom_line() +
scale_color_viridis_d(NULL, option = "A", end = .5) +
scale_x_continuous("grade", breaks = 6:12, limits = c(6, 12)) +
scale_y_continuous("estimated survival probability",
breaks = c(0, .5, 1), limits = c(0, 1)) +
theme(panel.grid = element_blank())
```

Combine the two **ggplot2** objects with **patchwork** syntax to make our version of Figure 11.1.

```
library(patchwork)
/ p2) +
(p1 plot_layout(guides = "collect")
```

On page 361, Singer and Willett compared the hazard probabilities at grades 8 and 11 for boys in the two `pt`

groups. We can make that comparison with `filter()`

.

```
%>%
lt filter(time %in% c(8, 11) &
!= "overall") %>%
pt select(pt:time, hazard_fun)
```

```
## # A tibble: 4 × 3
## pt time hazard_fun
## <fct> <dbl> <dbl>
## 1 pt = 0 8 0.0286
## 2 pt = 0 11 0.192
## 3 pt = 1 8 0.0526
## 4 pt = 1 11 0.283
```

Compare the two groups on the hazard probabilities at grade 9.

```
%>%
lt filter(time == 9 &
!= "overall") %>%
pt select(pt:time, hazard_fun)
```

```
## # A tibble: 2 × 3
## pt time hazard_fun
## <fct> <dbl> <dbl>
## 1 pt = 0 9 0.118
## 2 pt = 1 9 0.178
```

Now compare them on their hazard probabilities in grade 12.

```
%>%
lt filter(time == 12 &
!= "overall") %>%
pt select(pt:time, hazard_fun)
```

```
## # A tibble: 2 × 3
## pt time hazard_fun
## <fct> <dbl> <dbl>
## 1 pt = 0 12 0.190
## 2 pt = 1 12 0.474
```

At the top of page 362, Singer and Willett compared the percentages of boys who were virgins at grades 9 and 12, by `pt`

status. Those percentages are straight algebraic transformations of the corresponding survival function values.

```
%>%
lt filter(time %in% c(9, 12) &
!= "overall") %>%
pt select(pt:time, survivor_fun) %>%
mutate(percent_virgins = (100 * survivor_fun) %>% round(digits = 1))
```

```
## # A tibble: 4 × 4
## pt time survivor_fun percent_virgins
## <fct> <dbl> <dbl> <dbl>
## 1 pt = 0 9 0.833 83.3
## 2 pt = 0 12 0.472 47.2
## 3 pt = 1 9 0.685 68.5
## 4 pt = 1 12 0.185 18.5
```

Now let’s finish off by computing the interpolated median lifetime values for each with our custom `make_iml()`

function.

```
<- function(lt) {
make_iml
# lt is a generic name for a life table of the
# kind we made with our `make_lt()` function
# determine the mth row
<-
lt_m %>%
lt filter(survivor_fun > .5) %>%
slice(n())
# determine the row for m + 1
<-
lt_m1 %>%
lt filter(survivor_fun < .5) %>%
slice(1)
# pull the value for m
<- pull(lt_m, time)
m
# pull the two survival function values
<- pull(lt_m, survivor_fun)
stm <- pull(lt_m1, survivor_fun)
stm1
# plug the values into Equation 10.6 (page 338)
<- m + ((stm - .5) / (stm - stm1)) * ((m + 1) - m)
iml
iml
}
```

Put `make_iml()`

to work.

`make_iml(lt %>% filter(pt == "pt = 0"))`

`## [1] 11.75`

`make_iml(lt %>% filter(pt == "pt = 1"))`

`## [1] 9.952381`

### 11.1.2 What kind of statistical model do these graphs suggest?

To postulate a statistical model to represent the relationship between the population discrete-time hazard function and predictors, we must deal with two complications apparent in these displays. One is that any hypothesized model must describe the shape of the

entire discrete-time hazard functionover time, not just its value in any one period, in much the same way that a multilevel model for change characterizes the shape of entire individual growth trajectories over time. A second complication is that, as a conditional probability, the value of discrete-time hazard must lie between 0 and 1. Any reasonable statistical model for hazard must recognize this constraint, precluding the occurrence of theoretically impossible values. (p. 362,emphasisin the original)

#### 11.1.2.1 The bounded nature of hazard.

A conventional way to handle the bounded nature of probabilities is transform the scale of the data. David R. Cox (1972) recommended either the odds and log-odds (i.e., logit) transformations. Given a probability, \(p\), we compute the odds as

\[\text{odds} = \frac{p}{1 - p}.\]

Log-odds is a minor extension; you simply take the log of the odds, which we can formally express as

\[\text{log-odds} = \log \left (\frac{p}{1 - p} \right ).\]

To make the conversions easy, we’ll define^{2} a couple convenience functions: `odds()`

and `log_odds()`

.

```
<- function(p) {
odds / (1 - p)
p
}
<- function(p) {
log_odds log(p / (1 - p))
}
```

Here’s how they work.

```
tibble(p = seq(from = 0, to = 1, by = .1)) %>%
mutate(odds = odds(p),
log_odds = log_odds(p))
```

```
## # A tibble: 11 × 3
## p odds log_odds
## <dbl> <dbl> <dbl>
## 1 0 0 -Inf
## 2 0.1 0.111 -2.20
## 3 0.2 0.25 -1.39
## 4 0.3 0.429 -0.847
## 5 0.4 0.667 -0.405
## 6 0.5 1 0
## 7 0.6 1.5 0.405
## 8 0.7 2.33 0.847
## 9 0.8 4 1.39
## 10 0.9 9 2.20
## 11 1 Inf Inf
```

Before we make our version of Figure 11.2, it might be instructive to see how odds and log-odds compare to probabilities in a plot. Here we’ll compare them to probabilities ranging from .01 to .99.

```
tibble(p = seq(from = .01, to = .99, by = .01)) %>%
mutate(odds = odds(p),
`log(odds)` = log_odds(p)) %>%
pivot_longer(-p) %>%
mutate(name = factor(name,
levels = c("odds", "log(odds)"))) %>%
ggplot(aes(x = p, y = value)) +
geom_line() +
labs(x = "probability",
y = "transformed scale") +
theme(panel.grid = element_blank()) +
facet_wrap(~ name, scales = "free")
```

Odds are bounded to values of zero and above and have an inflection at 1. Log-odds are unbounded and have an inflection point at 0. Here we’ll save the odds and log-odds for our hazard functions within the `lt`

life table.

```
<-
lt %>%
lt mutate(odds = odds(hazard_fun),
log_odds = log_odds(hazard_fun))
lt
```

```
## # A tibble: 21 × 10
## pt time time_int n_risk n_event n_censored hazard_fun survivor_fun odds log_odds
## <fct> <dbl> <chr> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl>
## 1 pt = 0 6 [6, 7) 72 NA NA NA 1 NA NA
## 2 pt = 0 7 [7, 8) 72 2 0 0.0278 0.972 0.0286 -3.56
## 3 pt = 0 8 [8, 9) 70 2 0 0.0286 0.944 0.0294 -3.53
## 4 pt = 0 9 [9, 10) 68 8 0 0.118 0.833 0.133 -2.01
## 5 pt = 0 10 [10, 11) 60 8 0 0.133 0.722 0.154 -1.87
## 6 pt = 0 11 [11, 12) 52 10 0 0.192 0.583 0.238 -1.44
## 7 pt = 0 12 [12, 13) 42 8 34 0.190 0.472 0.235 -1.45
## 8 pt = 1 6 [6, 7) 108 NA NA NA 1 NA NA
## 9 pt = 1 7 [7, 8) 108 13 0 0.120 0.880 0.137 -1.99
## 10 pt = 1 8 [8, 9) 95 5 0 0.0526 0.833 0.0556 -2.89
## # ℹ 11 more rows
```

We’re ready to make and combine the subplots for our version of Figure 11.2.

```
# hazard
<-
p1 %>%
lt filter(pt != "overall") %>%
ggplot(aes(x = time, y = hazard_fun, color = pt, group = pt)) +
geom_line() +
scale_y_continuous(NULL, breaks = c(0, .5, 1), limits = c(0, 1)) +
labs(subtitle = "Estimated hazard") +
theme(legend.background = element_rect(fill = "transparent"),
legend.key = element_rect(color = "grey92"),
legend.position = c(.1, .825))
# odds
<-
p2 %>%
lt filter(pt != "overall") %>%
ggplot(aes(x = time, y = odds, color = pt, group = pt)) +
geom_line() +
scale_y_continuous(NULL, breaks = c(0, .5, 1), limits = c(0, 1)) +
labs(subtitle = "Estimated odds") +
theme(legend.position = "none")
# log-odds
<-
p3 %>%
lt filter(pt != "overall") %>%
ggplot(aes(x = time, y = log_odds, color = pt, group = pt)) +
geom_line() +
scale_y_continuous(NULL, limits = c(-4, 0)) +
labs(subtitle = "Estimated logit(hazard)") +
theme(legend.position = "none")
/ p2 / p3 ) &
(p1 scale_color_viridis_d(NULL, option = "A", end = .5) &
scale_x_continuous("grade", breaks = 6:12, limits = c(6, 12)) &
theme(panel.grid = element_blank())
```

#### 11.1.2.2 What statistical model could have generated these sample data?

With the survival models from the prior sections, we were lazy and just used the **survival** package. But recall from the end of the last chapter that we can fit the analogous models **brms** using the binomial likelihood. This subsection is a great place to practice those some more. The fitted lines Singer and Willett displayed in Figure 11.3 can all be reproduced with binomial regression. However, the `sex`

data are not in a convenient form to fit those models. Just like we did in last chapter, we’ll want to take a two-step process whereby we first convert the data to the long (i.e., person-period) format and then summarize. Happily, we can accomplish that first step by uploading the data in the `firstsex_pp.csv`

file, which are already in the long format.

```
<- read_csv("data/firstsex_pp.csv")
sex_pp
glimpse(sex_pp)
```

```
## Rows: 822
## Columns: 11
## $ id <dbl> 1, 1, 1, 2, 2, 2, 2, 2, 2, 3, 3, 3, 3, 3, 3, 5, 5, 5, 5, 5, 5, 6, 6, 6, 6, 6, 7, 7, 7, 9, 9, …
## $ period <dbl> 7, 8, 9, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 7, 8…
## $ event <dbl> 0, 0, 1, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, 0, 0, …
## $ d7 <dbl> 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, 0, …
## $ d8 <dbl> 0, 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, …
## $ d9 <dbl> 0, 0, 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, …
## $ d10 <dbl> 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, …
## $ d11 <dbl> 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, …
## $ d12 <dbl> 0, 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, …
## $ pt <dbl> 0, 0, 0, 1, 1, 1, 1, 1, 1, 0, 0, 0, 0, 0, 0, 1, 1, 1, 1, 1, 1, 0, 0, 0, 0, 0, 1, 1, 1, 0, 0, …
## $ pas <dbl> 1.9788670, 1.9788670, 1.9788670, -0.5454916, -0.5454916, -0.5454916, -0.5454916, -0.5454916, …
```

Now we’ll compute the desired summary values and wrangle a bit.

```
<-
sex_aggregated %>%
sex_pp mutate(event = if_else(event == 1, "event", "no_event")) %>%
group_by(period) %>%
count(event, pt) %>%
ungroup() %>%
pivot_wider(names_from = event,
values_from = n) %>%
mutate(total = event + no_event,
period_center = period - mean(period),
peroid_factor = factor(period),
pt = factor(pt))
sex_aggregated
```

```
## # A tibble: 12 × 7
## period pt event no_event total period_center peroid_factor
## <dbl> <fct> <int> <int> <int> <dbl> <fct>
## 1 7 0 2 70 72 -2.5 7
## 2 7 1 13 95 108 -2.5 7
## 3 8 0 2 68 70 -1.5 8
## 4 8 1 5 90 95 -1.5 8
## 5 9 0 8 60 68 -0.5 9
## 6 9 1 16 74 90 -0.5 9
## 7 10 0 8 52 60 0.5 10
## 8 10 1 21 53 74 0.5 10
## 9 11 0 10 42 52 1.5 11
## 10 11 1 15 38 53 1.5 11
## 11 12 0 8 34 42 2.5 12
## 12 12 1 18 20 38 2.5 12
```

Note how we saved the grade values in three columns:

`period`

has them as continuous values, which will be hand for plotting;`period_center`

has them as mean-centered continuous values, which will make fitting the linear model in the middle panel easier; and`period_factor`

has them saved as a factor, which will help us fit the model in the bottom panel.

Fire up **brms**.

`library(brms)`

Before we fit the models, it might be good to acknowledge we’re jumping ahead of the authors, a bit. Singer and Willett didn’t discuss fitting discrete time hazard models until section 11.3.2. Sure, their focus was on the frequentist approach using maximum likelihood. But the point still stands. If these model fitting details feel a bit rushed, they are.

Any anxious feelings aside, now fit the three binomial models. We continue to use weakly-regularizing priors for each.

```
# top panel
.4 <-
fit11brm(data = sex_aggregated,
family = binomial,
| trials(total) ~ 0 + pt,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.04")
# middle panel
.5 <-
fit11brm(data = sex_aggregated,
family = binomial,
| trials(total) ~ 0 + pt + period_center,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.05")
# bottom panel
.6 <-
fit11brm(data = sex_aggregated,
family = binomial,
| trials(total) ~ 0 + pt + peroid_factor,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.06")
```

Check the model summaries.

`print(fit11.4)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(total) ~ 0 + pt
## Data: sex_aggregated (Number of observations: 12)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## pt0 -2.15 0.17 -2.50 -1.83 1.00 3668 2824
## pt1 -1.44 0.12 -1.68 -1.21 1.00 4237 2963
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

`print(fit11.5)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(total) ~ 0 + pt + period_center
## Data: sex_aggregated (Number of observations: 12)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## pt0 -2.23 0.18 -2.60 -1.89 1.00 3779 3003
## pt1 -1.35 0.12 -1.60 -1.12 1.00 3270 2585
## period_center 0.43 0.06 0.31 0.56 1.00 3635 3089
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

`print(fit11.6)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(total) ~ 0 + pt + peroid_factor
## Data: sex_aggregated (Number of observations: 12)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## pt0 -3.00 0.32 -3.63 -2.38 1.00 986 1718
## pt1 -2.12 0.27 -2.68 -1.61 1.00 912 1224
## peroid_factor8 -0.77 0.48 -1.76 0.12 1.00 1442 1993
## peroid_factor9 0.69 0.35 0.03 1.37 1.00 1096 1527
## peroid_factor10 1.15 0.34 0.50 1.83 1.00 1064 1732
## peroid_factor11 1.32 0.36 0.61 2.03 1.00 1142 1581
## peroid_factor12 1.80 0.36 1.11 2.50 1.00 1180 1759
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

We can extract the fitted values and their summaries for each row in the data with `fitted()`

. To get them in the log-odds metric, we need to set `scale = "linear"`

. Here’s a quick example with `fit11.4`

.

`fitted(fit11.4, scale = "linear")`

```
## Estimate Est.Error Q2.5 Q97.5
## [1,] -2.153394 0.1698095 -2.501172 -1.829720
## [2,] -1.440906 0.1215326 -1.682504 -1.210137
## [3,] -2.153394 0.1698095 -2.501172 -1.829720
## [4,] -1.440906 0.1215326 -1.682504 -1.210137
## [5,] -2.153394 0.1698095 -2.501172 -1.829720
## [6,] -1.440906 0.1215326 -1.682504 -1.210137
## [7,] -2.153394 0.1698095 -2.501172 -1.829720
## [8,] -1.440906 0.1215326 -1.682504 -1.210137
## [9,] -2.153394 0.1698095 -2.501172 -1.829720
## [10,] -1.440906 0.1215326 -1.682504 -1.210137
## [11,] -2.153394 0.1698095 -2.501172 -1.829720
## [12,] -1.440906 0.1215326 -1.682504 -1.210137
```

If we convert that output to a data frame, tack on the original data values, and wrangle a bit, we’ll be in good shape to make the top panel of Figure 11.3. Below we’ll do that for each of the three panels, saving them as `p1`

, `p2`

, and `p3`

.

```
# logit(hazard) is horizontal with time
<-
p1 fitted(fit11.4,
scale = "linear") %>%
data.frame() %>%
bind_cols(sex_aggregated) %>%
mutate(pt = str_c("pt = ", pt)) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/4) +
geom_line(aes(y = Estimate),
alpha = 1/2) +
geom_point(aes(y = log_odds(event / total))) +
scale_y_continuous(NULL, limits = c(-4, 0)) +
labs(subtitle = "logit(hazard) is horizontal with time") +
theme(legend.background = element_rect(fill = "transparent"),
legend.key = element_rect(color = "grey92"),
legend.position = c(.1, .825))
# logit(hazard) is linear with time
<-
p2 fitted(fit11.5,
scale = "linear") %>%
data.frame() %>%
bind_cols(sex_aggregated) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/4) +
geom_line(aes(y = Estimate),
alpha = 1/2) +
geom_point(aes(y = log_odds(event / total))) +
labs(subtitle = "logit(hazard) is linear with time",
y = "logit(hazard)") +
coord_cartesian(ylim = c(-4, 0)) +
theme(legend.position = "none")
# logit(hazard) is completely general with time
<-
p3 fitted(fit11.6,
scale = "linear") %>%
data.frame() %>%
bind_cols(sex_aggregated) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/4) +
geom_line(aes(y = Estimate),
alpha = 1/2) +
geom_point(aes(y = log_odds(event / total))) +
labs(subtitle = "logit(hazard) is completely general with time",
y = NULL) +
coord_cartesian(ylim = c(-4, 0)) +
theme(legend.position = "none")
```

Now combine the plots with **patchwork** syntax.

```
/ p2 / p3) &
(p1 scale_fill_viridis_d(NULL, option = "A", end = .6) &
scale_color_viridis_d(NULL, option = "A", end = .6) &
scale_x_continuous("Grade", breaks = 6:12, limits = c(6, 12)) &
theme(panel.grid = element_blank())
```

In addition to the posterior means (i.e., our analogues to the fitted values in the text), we added the 95% Bayesian intervals to give a better sense of the uncertainty in each model. Singer and Willett mused the unconstrained model (`fit6`

) was a better fit to the data than the other two. We can quantify that with a LOO comparison.

```
.4 <- add_criterion(fit11.4, "loo")
fit11.5 <- add_criterion(fit11.5, "loo")
fit11.6 <- add_criterion(fit11.6, "loo")
fit11
loo_compare(fit11.4, fit11.5, fit11.6) %>% print(simplify = F)
```

```
## elpd_diff se_diff elpd_loo se_elpd_loo p_loo se_p_loo looic se_looic
## fit11.6 0.0 0.0 -31.6 1.7 5.1 0.7 63.1 3.5
## fit11.5 -0.7 3.2 -32.2 3.2 3.5 1.2 64.5 6.4
## fit11.4 -27.7 10.5 -59.3 10.4 9.3 2.6 118.6 20.7
```

Here are the LOO weights.

`model_weights(fit11.4, fit11.5, fit11.6, weights = "loo") %>% round(digits = 3)`

```
## fit11.4 fit11.5 fit11.6
## 0.000 0.338 0.662
```

## 11.2 Formal representation of the population discrete-time hazard model

Earlier equations for the hazard function omitted substantive predictors. Now consider the case where \(X_{1ij}, X_{2ij}, \dots , X_{Pij}\) stand for the \(P\) predictors which may or may not vary across individuals \(i\) and time periods \(j\). Thus \(x_{pij}\) is the value for the \(i^\text{th}\) individual on the \(p^\text{th}\) variable during the \(j^\text{th}\) period. We can use this to define the conditional hazard function as

\[h(t_{ij}) = \Pr [T_i = j \mid T \geq j \text{ and } X_{1ij} = x_{1ij}, X_{2ij} = x_{2ij}, \dots , X_{Pij} = x_{pij}].\]

Building further and keeping the baseline shape of the discrete hazard function flexible, we want a method that allows each of the \(j\) time periods to have its own value. Imagine a set of \(J\) dummy variables, \(D_1, D_2, \dots, D_J\), marking off each of the time periods. For example, say \(J = 6\), we could depict this in a tibble like so.

```
tibble(period = 1:6) %>%
mutate(d1 = if_else(period == 1, 1, 0),
d2 = if_else(period == 2, 1, 0),
d3 = if_else(period == 3, 1, 0),
d4 = if_else(period == 4, 1, 0),
d5 = if_else(period == 5, 1, 0),
d6 = if_else(period == 6, 1, 0))
```

```
## # A tibble: 6 × 7
## period d1 d2 d3 d4 d5 d6
## <int> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl>
## 1 1 1 0 0 0 0 0
## 2 2 0 1 0 0 0 0
## 3 3 0 0 1 0 0 0
## 4 4 0 0 0 1 0 0
## 5 5 0 0 0 0 1 0
## 6 6 0 0 0 0 0 1
```

If we were to use a set of dummies of this kind in a model, we would omit the conventional regression intercept, replacing it with the \(J\) dummies. Now presume we’re fitting a hazard model using the logit link, \(\operatorname{logit} h(t_{ij})\). We can express the discrete conditional hazard model with a general functional form with respect to time as

\[\operatorname{logit} h(t_{ij}) = [\alpha_1 D_{1ij} + \alpha_2 D_{2ij} + \cdots + \alpha_J D_{Jij}] + [\beta_1 X_{1ij} + \beta_2 X_{2ij} + \cdots + \beta_P X_{Pij}],\]

where the \(\alpha\) parameters are the \(J\) time-period dummies and the \(\beta\) parameters are for other time-varying or time-invariant predictors. This is just the type of model we used to fit `fit116`

. For that model, the basic equation was

\[\operatorname{logit} h(t_{ij}) = [\alpha_7 D_{7ij} + \alpha_8 D_{8ij} + \cdots + \alpha_{12} D_{12ij}] + [\beta_1 \text{PT}_i ],\]

where the only substantive predictor was the time-invariant `pt`

. However, that formula could be a little misleading. Recall the formula:

`.6$formula fit11`

`## event | trials(total) ~ 0 + pt + peroid_factor`

We suppressed the default regression intercept with the `~ 0 +`

syntax and the only two predictors were `pt`

and `peroid_factor`

. Both were saved as factor variables. Functionally, that’s why `period_factor`

worked as \(\alpha_7 D_{7ij} + \alpha_8 D_{8ij} + \cdots + \alpha_{12} D_{12ij}\), a series of 5 dummy variables with no reference category. The same basic thing goes for `pt`

. Because `pt`

was a factor used in a model `formula`

with no conventional intercept, it acted as if it was a series of 2 dummy variables with no reference category. Thus, we might rewrite the model equation for `fit6`

as

\[\operatorname{logit} h(t_{ij}) = [\alpha_7 D_{7ij} + \alpha_8 D_{8ij} + \cdots + \alpha_{12} D_{12ij}] + [\beta_0 \text{PT}_{0i} + \beta_1 \text{PT}_{1i} ].\]

And since we’re practicing fitting these models as Bayesians, the `fit6`

equation with a fuller expression of the likelihood and the priors looks like

\[ \begin{align*} \text{event}_{ij} & = \operatorname{Binomial}(n = \text{trials}_{ij}, p_{ij}) \\ \operatorname{logit} (p_{ij}) & = [\alpha_7 D_{7ij} + \alpha_8 D_{8ij} + \cdots + \alpha_{12} D_{12ij}] + [\beta_0 \text{PT}_{0i} + \beta_1 \text{PT}_{1i} ] \\ \alpha_7, \alpha_8, \dots, \alpha_{12} & \sim \operatorname{Normal}(0, 4) \\ \beta_0 \text{ and } \beta_1 & \sim \operatorname{Normal}(0, 4), \end{align*} \]

where we’re describing the model in terms of the criterion, `event`

, rather than in terms of \(h(t_{ij})\). And what is the criterion, `event`

? It’s a vector of counts. The binomial likelihood allows us to model vectors of counts in terms of the number of trials, as indexed by our `trials`

vector, and the (conditional) probability of a “1” in a given trial. In this context, \(h(t_{ij}) = p_{ij}\).

### 11.2.1 What do the parameters represent?

Given our factor coding of `pt`

, our two submodels for the equations in the last section are

\[ \begin{align*} \text{when PT = 0: } \operatorname{logit} h(t_j) & = [\alpha_7 D_7 + \alpha_8 D_8 + \cdots + \alpha_{12} D_{12}] + \beta_0 \\ \text{when PT = 1: } \operatorname{logit} h(t_j) & = [\alpha_7 D_7 + \alpha_8 D_8 + \cdots + \alpha_{12} D_{12}] + \beta_1, \end{align*} \]

where we used Singer and Willett’s simplified notation and dropped all the \(i\) subscripts and most of the \(j\) subscripts.

### 11.2.2 An alternative representation of the model.

In the previous sections, we expressed the model in terms of the logit of the criterion or the \(p\) parameter of the likelihood. Another strategy is the express the criterion (or \(p\)) in its natural metric and put the nonlinear portion on the right side of the equation. If we consider the generic discrete conditional hazard function, that would follow the form

\[ h(t_{ij}) = \frac{1}{1 + e^{-([\alpha_1 D_{1ij} + \alpha_2 D_{2ij} + \cdots + \alpha_J D_{Jij}] + [\beta_1 X_{1ij} + \beta_2 X_{2ij} + \cdots + \beta_P X_{Pij}])}}. \]

This is just a particular kind of logistic regression model. It also clarifies that “by specifying a linear relationship between predictors and logit hazard we imply a *nonlinear* relationship between predictors and *raw hazard*” (p. 377, *emphasis* in the original). We can explore what that might look like with our version of Figure 11.4. Here we continue to use `fit6`

, but this time we’ll save the output from `fitted()`

before plotting.

```
<-
f fitted(fit11.6,
scale = "linear") %>%
data.frame() %>%
bind_cols(sex_aggregated)
f
```

```
## Estimate Est.Error Q2.5 Q97.5 period pt event no_event total period_center peroid_factor
## 1 -2.9970126 0.3176158 -3.6294093 -2.3832815 7 0 2 70 72 -2.5 7
## 2 -2.1180003 0.2735893 -2.6848898 -1.6111410 7 1 13 95 108 -2.5 7
## 3 -3.7684632 0.4287806 -4.6415348 -2.9903587 8 0 2 68 70 -1.5 8
## 4 -2.8894508 0.3951568 -3.7108161 -2.1724589 8 1 5 90 95 -1.5 8
## 5 -2.3082801 0.2743845 -2.8655483 -1.7835310 9 0 8 60 68 -0.5 9
## 6 -1.4292677 0.2334189 -1.9071366 -0.9685727 9 1 16 74 90 -0.5 9
## 7 -1.8435758 0.2576237 -2.3635566 -1.3492209 10 0 8 52 60 0.5 10
## 8 -0.9645635 0.2215530 -1.4188340 -0.5291742 10 1 21 53 74 0.5 10
## 9 -1.6761430 0.2710102 -2.2258520 -1.1397280 11 0 10 42 52 1.5 11
## 10 -0.7971307 0.2476662 -1.2959928 -0.3393064 11 1 15 38 53 1.5 11
## 11 -1.2014513 0.2681393 -1.7501715 -0.6750968 12 0 8 34 42 2.5 12
## 12 -0.3224389 0.2584378 -0.8478826 0.1663958 12 1 18 20 38 2.5 12
```

Make the subplots.

```
# logit(hazard)
<-
p1 %>%
f mutate(pt = str_c("pt = ", pt)) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "logit(hazard)",
y = NULL) +
coord_cartesian(ylim = c(-4, 0)) +
theme(legend.background = element_rect(fill = "transparent"),
legend.key = element_rect(color = "grey92"),
legend.position = c(.1, .825))
# odds
<-
p2 %>%
f mutate_at(vars(Estimate, Q2.5, Q97.5), .funs = exp) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "odds",
y = NULL) +
coord_cartesian(ylim = c(0, .8)) +
theme(legend.position = "none")
# hazard
<-
p3 %>%
f mutate_at(vars(Estimate, Q2.5, Q97.5), .funs = inv_logit_scaled) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "hazard (i.e., probability)",
y = NULL) +
coord_cartesian(ylim = c(0, .5)) +
theme(legend.position = "none")
```

Combine the subplots with **patchwork**.

```
/ p2 / p3) &
(p1 scale_fill_viridis_d(NULL, option = "A", end = .6) &
scale_color_viridis_d(NULL, option = "A", end = .6) &
scale_x_continuous("Grade", breaks = 6:12, limits = c(6, 12)) &
theme(panel.grid = element_blank())
```

The `mutate_at()`

conversions we made for `p2`

and `p3`

were based on the guidelines in Table 11.2. Those were:

```
tibble(`original scale` = c("logit", "odds", "logit"),
`desired scale` = c("odds", "probability", "probability"),
transformation = c("exp(logit)", "odds / (1 + odds)", "1 / (1 + exp(-1 * logit))")) %>%
::flextable() %>%
flextable::width(width = c(1.5, 1.5, 2)) flextable
```

original scale | desired scale | transformation |
---|---|---|

logit | odds | exp(logit) |

odds | probability | odds / (1 + odds) |

logit | probability | 1 / (1 + exp(-1 * logit)) |

We accomplished the transformation in the bottom row with the `brms::inv_logit_scaled()`

function.

## 11.3 Fitting a discrete-time hazard model to data

As Singer and Willett wrote, “with data collected on a random sample of individuals from a target population, you can easily fit a discrete-time hazard model, estimate its parameters using maximum likelihood methods, and evaluate goodness-of-fit” (pp. 378–379. As we’ve already demonstrated, you can fit them with Bayesian software, too. Though we’ll be focusing on **brms**, you might also want to check out the **rstanarm** package, about which you can learn more from Brilleman, Elci, Novik, and Wolfe’s (2020) preprint, *Bayesian survival analysis Using the rstanarm R package*, Brilleman’s (2019) vignette, *Estimating survival (time-to-event) models with rstanarm*, and the Survival models in rstanarm thread in the Stan forums.

### 11.3.1 Adding predictors to the person-period data set.

At the beginning of section 11.1.2.2, we already loaded a version of the person-period data including the discrete-time dummies. It has our substantive predictors `pt`

and `pas`

, too. We saved it as `sex_pp`

. Here’s a `glimpse()`

.

```
%>%
sex_pp glimpse()
```

```
## Rows: 822
## Columns: 11
## $ id <dbl> 1, 1, 1, 2, 2, 2, 2, 2, 2, 3, 3, 3, 3, 3, 3, 5, 5, 5, 5, 5, 5, 6, 6, 6, 6, 6, 7, 7, 7, 9, 9, …
## $ period <dbl> 7, 8, 9, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 12, 7, 8, 9, 10, 11, 7, 8…
## $ event <dbl> 0, 0, 1, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, 0, 0, …
## $ d7 <dbl> 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, 0, …
## $ d8 <dbl> 0, 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, 1, …
## $ d9 <dbl> 0, 0, 1, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 1, 0, 0, …
## $ d10 <dbl> 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, …
## $ d11 <dbl> 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, …
## $ d12 <dbl> 0, 0, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 0, 0, 0, 0, …
## $ pt <dbl> 0, 0, 0, 1, 1, 1, 1, 1, 1, 0, 0, 0, 0, 0, 0, 1, 1, 1, 1, 1, 1, 0, 0, 0, 0, 0, 1, 1, 1, 0, 0, …
## $ pas <dbl> 1.9788670, 1.9788670, 1.9788670, -0.5454916, -0.5454916, -0.5454916, -0.5454916, -0.5454916, …
```

### 11.3.2 Maximum likelihood estimates [and Bayesian posteriors] for the discrete-time hazard model.

We’re not going to walk through all the foundational equations for the likelihood and log-likelihood functions (Equations 11.11 through 11.13). For our purposes, just note that “it turns out that the standard logistic regression routines widely available in all major statistical packages, *when applied appropriately in the person-period data set*, actually provide estimates of the parameters of the discrete-time hazard model” (p. 383, *emphasis* in the original). Happily, this is what we’ve been doing. Bayesian logistic regression via the binomial likelihood has been our approach. And since we’re Bayesians, the same caveat applies to survival models as applied to the other longitudinal models we fit in earlier chapters. We’re not just maximizing likelihoods, here. Bayes’s formula requires us to multiply the likelihood by the prior.

\[ \\underbrace{p(\theta \mid d)}_\text{posterior} \propto \underbrace{p(d \mid \theta)}_\text{likelihood} \; \underbrace{p(\theta)}_\text{prior} \]

### 11.3.3 Fitting the discrete-time hazard model to data.

In one sense, fitting discrete-hazard models with Bayesian logistic regression is old hat, for us. We’ve been doing that since the end of last chapter. But one thing I haven’t clarified is, up to this point, we have been using the aggregated binomial format. To show what I mean, we might look at the data we used for our last model, `fit11.6`

.

` sex_aggregated`

```
## # A tibble: 12 × 7
## period pt event no_event total period_center peroid_factor
## <dbl> <fct> <int> <int> <int> <dbl> <fct>
## 1 7 0 2 70 72 -2.5 7
## 2 7 1 13 95 108 -2.5 7
## 3 8 0 2 68 70 -1.5 8
## 4 8 1 5 90 95 -1.5 8
## 5 9 0 8 60 68 -0.5 9
## 6 9 1 16 74 90 -0.5 9
## 7 10 0 8 52 60 0.5 10
## 8 10 1 21 53 74 0.5 10
## 9 11 0 10 42 52 1.5 11
## 10 11 1 15 38 53 1.5 11
## 11 12 0 8 34 42 2.5 12
## 12 12 1 18 20 38 2.5 12
```

Now recall the formula for the binomial likelihood from the end of last chapter:

\[\Pr (z \mid n, p) = \frac{n!}{z!(n - z)!} p^z (1 - p)^{n - z},\]

where \(z\) is the number of cases for which the value is 1, \(n\) is the total number of cases, and \(p\) is the constant chance of a 1 across cases. We refer to binomial data as aggregated with \(n > 1\). Our \(n\) vector in the `sex_aggregated`

, `total`

, ranged from 38 to 108. Accordingly, our \(z\) vector, `event`

, was always some value equal or lower to that in the same row for `total`

.

The person-period data, `sex_pp`

, contain the same information but in a different format. Instead, each `event`

cell only takes on a value of 0 or 1 (i.e., \(n = 1\)). If you were to sum up all the values in the `total`

column of the `sex_aggregated`

data, you’d return 822.

```
%>%
sex_aggregated summarise(sum = sum(total))
```

```
## # A tibble: 1 × 1
## sum
## <int>
## 1 822
```

This is also the total number of rows in the `sex_pp`

data.

```
%>%
sex_pp count()
```

```
## # A tibble: 1 × 1
## n
## <int>
## 1 822
```

It’s also the case that when \(n = 1\), the right side of the equation for the binomial function reduces to

\[p^z (1 - p)^{1 - z}.\]

Whether you are working with aggregated or un-aggregated data, both are suited to fit logistic regression models with the binomial likelihood. Just specify the necessary information in the model syntax. For **brms**, the primary difference is how you use the `trials()`

function. When we fit our logistic regression models using the aggregated data, we specified `trials(total)`

, which informed the `brm()`

function what the \(n\) values were. In the case of unaggregated binomial data, we can just state `trials(1)`

. Each cell is the outcome \(z\) for a single trial.

Before we fit the models, we might talk a bit about priors. When we fit the first model of this kind at the end of Chapter 10, we just used `prior(normal(0, 4), class = b)`

without comment. Recall we’re modeling probabilities in the log-odds space. In Section 11.1.2.1 we used a plot to compare probability values to their log-odds counterparts. Let’s take a more focused look.

```
tibble(log_odds = -8:8) %>%
mutate(p = inv_logit_scaled(log_odds)) %>%
ggplot(aes(x = log_odds, y = p)) +
geom_hline(yintercept = 0:5 / 5, color = "white") +
geom_point() +
scale_x_continuous(breaks = -8:8) +
scale_y_continuous(breaks = 0:5 / 5) +
theme(panel.grid = element_blank())
```

When \(\operatorname{log-odds} p = 0\), \(p = .5\). Once \(\operatorname{log-odds} p\) approaches the \(\mp 4\) neighborhood, the corresponding values for \(p\) asymptote at the boundaries \([0, 1]\). By using a \(\operatorname{Normal} (0, 4)\) prior for \(\operatorname{log-odds} p\), we’re putting bulk of the prior mass in the \(\operatorname{log-odds} p\) space between, say, -8 and 8. In the absence of other information, this might be a good place to start. A little further down, we’ll reexamine this set-up. For now, here’s how to use `brm()`

to fit Models A through D from page 385.

```
library(brms)
# model a
.7 <-
fit11brm(data = sex_pp,
family = binomial,
| trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.07")
# model b
.8 <-
fit11brm(data = sex_pp,
family = binomial,
| trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.08")
# model c
.9 <-
fit11brm(data = sex_pp,
family = binomial,
| trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pas,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.09")
# model d
.10 <-
fit11brm(data = sex_pp,
family = binomial,
| trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt + pas,
event prior(normal(0, 4), class = b),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.10")
```

## 11.4 Interpreting parameter estimates

Here are the model summaries in bulk.

`print(fit11.7)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -2.43 0.27 -2.99 -1.92 1.00 6107 3046
## d8 -3.15 0.39 -3.96 -2.45 1.00 5698 3058
## d9 -1.73 0.23 -2.19 -1.29 1.00 6071 3032
## d10 -1.30 0.21 -1.72 -0.90 1.00 5271 2893
## d11 -1.18 0.24 -1.64 -0.73 1.00 5793 3056
## d12 -0.74 0.24 -1.20 -0.28 1.00 6444 3114
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

`print(fit11.8)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -3.02 0.32 -3.65 -2.41 1.00 3073 2485
## d8 -3.73 0.43 -4.66 -2.95 1.00 3922 2734
## d9 -2.29 0.28 -2.87 -1.78 1.00 2565 2672
## d10 -1.83 0.26 -2.34 -1.34 1.00 2747 2757
## d11 -1.65 0.27 -2.21 -1.14 1.00 3296 2575
## d12 -1.18 0.27 -1.72 -0.67 1.00 3041 2491
## pt 0.86 0.22 0.43 1.29 1.00 1908 2457
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

`print(fit11.9)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pas
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -2.49 0.28 -3.05 -1.97 1.00 6497 3093
## d8 -3.19 0.39 -4.02 -2.48 1.00 6204 2791
## d9 -1.75 0.23 -2.22 -1.32 1.00 6580 2693
## d10 -1.30 0.21 -1.71 -0.90 1.00 6685 2950
## d11 -1.15 0.23 -1.61 -0.71 1.00 7435 2591
## d12 -0.64 0.24 -1.14 -0.17 1.00 7176 3290
## pas 0.45 0.11 0.22 0.66 1.00 7266 3341
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

`print(fit11.10)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt + pas
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -2.91 0.33 -3.58 -2.30 1.00 2602 2358
## d8 -3.61 0.42 -4.47 -2.86 1.00 3035 2921
## d9 -2.15 0.28 -2.72 -1.63 1.00 2148 2448
## d10 -1.69 0.27 -2.23 -1.17 1.00 2187 2907
## d11 -1.52 0.28 -2.11 -0.98 1.00 2639 2409
## d12 -1.01 0.28 -1.58 -0.45 1.00 2571 2681
## pt 0.64 0.24 0.16 1.11 1.01 1358 1532
## pas 0.30 0.12 0.06 0.54 1.00 2849 2810
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

Although the text distinguishes between \(\alpha\) and \(\beta\) parameters (i.e., intercept and slope parameters, respectively), our **brms** output makes no such distinction. These are all of `class = b`

, population-level \(\beta\) parameters.

When viewed in bulk, all those `print()`

calls yield a lot of output. We can arrange the parameter summaries similar to those in Table 11.3 with a little tricky wrangling.

```
tibble(model = str_c("model ", letters[1:4]),
fit = str_c("fit11.", 7:10)) %>%
mutate(f = map(fit, ~ get(.) %>%
fixef() %>%
data.frame() %>%
rownames_to_column("parameter"))) %>%
unnest(f) %>%
mutate(e_sd = str_c(round(Estimate, digits = 2), " (", round(Est.Error, digits = 2), ")")) %>%
select(model, parameter, e_sd) %>%
pivot_wider(names_from = model, values_from = e_sd) %>%
::flextable() %>%
flextable::width(width = 1) flextable
```

parameter | model a | model b | model c | model d |
---|---|---|---|---|

d7 | -2.43 (0.27) | -3.02 (0.32) | -2.49 (0.28) | -2.91 (0.33) |

d8 | -3.15 (0.39) | -3.73 (0.43) | -3.19 (0.39) | -3.61 (0.42) |

d9 | -1.73 (0.23) | -2.29 (0.28) | -1.75 (0.23) | -2.15 (0.28) |

d10 | -1.3 (0.21) | -1.83 (0.26) | -1.3 (0.21) | -1.69 (0.27) |

d11 | -1.18 (0.24) | -1.65 (0.27) | -1.15 (0.23) | -1.52 (0.28) |

d12 | -0.74 (0.24) | -1.18 (0.27) | -0.64 (0.24) | -1.01 (0.28) |

pt | 0.86 (0.22) | 0.64 (0.24) | ||

pas | 0.45 (0.11) | 0.3 (0.12) |

### 11.4.1 The time indicators.

As a group, the \(\hat \alpha\)s are [Bayesian] estimates for the baseline logit hazard function. The amount and direction of variation in their values describe the shape of this function and tell us whether risk increases, decreases, or remains steady over time. (p. 387)

A coefficient plot might help us get a sense of that across the four models.

```
tibble(model = str_c("model ", letters[1:4]),
fit = str_c("fit11.", 7:10)) %>%
mutate(f = map(fit, ~ get(.) %>%
fixef() %>%
data.frame() %>%
rownames_to_column("parameter"))) %>%
unnest(f) %>%
filter(str_detect(parameter, "d")) %>%
mutate(parameter = factor(str_remove(parameter, "b_"),
levels = str_c("d", 12:7))) %>%
ggplot(aes(x = Estimate, xmin = Q2.5, xmax = Q97.5, y = parameter)) +
geom_pointrange(fatten = 2.5) +
labs(x = "posterior (log-odds scale)",
y = NULL) +
theme(axis.text.y = element_text(hjust = 0),
axis.ticks.y = element_blank(),
panel.grid = element_blank()) +
facet_wrap(~ model, nrow = 1)
```

“The fairly steady increase over time in the magnitude of the \(\hat \alpha\)s in each model [in the coefficient plot] shows that, in this sample of boys, the risk of first intercourse increases over time” (p. 387). When comparing the \(\hat \alpha\)s across models, it’s important to recall that the presence/absence of substantive covariates means each model has a different baseline group.

Because they were in the log-odds scale, the model output and our coefficient plot can be difficult to interpret. With the `brms::inv_logit_scaled()`

, we can convert the \(\hat \alpha\)s to the hazard (i.e., probability) metric.

```
tibble(model = str_c("model ", letters[1:4]),
fit = str_c("fit11.", 7:10)) %>%
mutate(f = map(fit, ~ get(.) %>%
fixef() %>%
data.frame() %>%
rownames_to_column("parameter"))) %>%
unnest(f) %>%
filter(str_detect(parameter, "d")) %>%
mutate(parameter = factor(str_remove(parameter, "b_"),
levels = str_c("d", 12:7))) %>%
mutate_at(vars(Estimate:Q97.5), .funs = inv_logit_scaled) %>%
ggplot(aes(x = Estimate, xmin = Q2.5, xmax = Q97.5, y = parameter)) +
geom_pointrange(fatten = 2.5) +
labs(x = "posterior (hazard scale)",
y = NULL) +
theme(axis.text.y = element_text(hjust = 0),
axis.ticks.y = element_blank(),
panel.grid = element_blank()) +
facet_wrap(~ model, nrow = 1)
```

Building further, here’s our version of Table 11.4.

```
fixef(fit11.7) %>%
data.frame() %>%
rownames_to_column("predictor") %>%
mutate(`time period` = str_remove(predictor, "d") %>% as.double()) %>%
select(`time period`, predictor, Estimate) %>%
mutate(`fitted odds` = exp(Estimate),
`fitted hazard` = inv_logit_scaled(Estimate)) %>%
mutate_if(is.double, round, digits = 4) %>%
::flextable() %>%
flextable::width(width = 1) flextable
```

time period | predictor | Estimate | fitted odds | fitted hazard |
---|---|---|---|---|

7 | d7 | -2.4253 | 0.0884 | 0.0813 |

8 | d8 | -3.1504 | 0.0428 | 0.0411 |

9 | d9 | -1.7294 | 0.1774 | 0.1507 |

10 | d10 | -1.2973 | 0.2733 | 0.2146 |

11 | d11 | -1.1759 | 0.3086 | 0.2358 |

12 | d12 | -0.7419 | 0.4762 | 0.3226 |

### 11.4.2 Dichotomous substantive predictors.

Here’s the summary for `pt`

from `fit11.8`

(i.e., Model B).

`fixef(fit11.8)["pt", ]`

```
## Estimate Est.Error Q2.5 Q97.5
## 0.8631546 0.2188621 0.4319023 1.2944748
```

If we take the anti-log (i.e., exponentiate) of that coefficient, we’ll return an odds ratio. Here’s the conversion with just the posterior mean.

`fixef(fit11.8)["pt", 1] %>% exp()`

`## [1] 2.370627`

To get a better sense of the conversion, here it is in a plot.

```
library(tidybayes)
posterior_samples(fit11.8) %>%
transmute(`log-odds` = b_pt,
`hazard ratio` = exp(b_pt)) %>%
pivot_longer(everything()) %>%
mutate(name = factor(name, levels = c("log-odds", "hazard ratio"))) %>%
ggplot(aes(x = value, y = 0)) +
stat_halfeye(.width = c(.5, .95), normalize = "panels") +
scale_y_continuous(NULL, breaks = NULL) +
xlab("marginal posterior for pt") +
theme(panel.grid = element_blank()) +
facet_wrap(~ name, scales = "free")
```

This tells us that, in every grade, the estimated odds of first intercourse are nearly two and one half times higher for boys who experienced a parenting transition in comparison to boys raised with both biological parents. In substantive terms, an odds ratio of this magnitude represents a substantial, and potentially important, effect. (p. 398)

To reframe the odds ratio in terms of the other group (i.e., `pt == 0`

), take the reciprocal.

`1 / exp(fixef(fit11.8)[7, 1])`

`## [1] 0.4218293`

“This tells us that the estimated odds of first intercourse for boys who did not experience a parenting transition are approximately 40% of the odds for boys who did. These complimentary ways of reporting effect sizes are equivalent” (p. 389)

### 11.4.3 Continuous substantive predictors.

Here’s the conditional effect of `pas`

from `fit11.9`

(i.e., Model C).

`fixef(fit11.9)["pas", ]`

```
## Estimate Est.Error Q2.5 Q97.5
## 0.4451985 0.1128906 0.2244570 0.6648630
```

To understand `pas`

, our measure of parental antisocial behavior, it will help to look at its range.

`range(sex_pp$pas)`

`## [1] -1.716180 2.781413`

Exponentiating (i.e., taking the anti-log) the posterior of a continuous predictor is a legitimate way to convert it to a hazard ratio.

```
as_draws_df(fit11.9) %>%
transmute(`log-odds` = b_pas,
`hazard ratio` = exp(b_pas)) %>%
pivot_longer(everything()) %>%
mutate(name = factor(name, levels = c("log-odds", "hazard ratio"))) %>%
ggplot(aes(x = value, y = 0)) +
stat_halfeye(.width = c(.5, .95), normalize = "panels") +
scale_y_continuous(NULL, breaks = NULL) +
xlab("marginal posterior for pas") +
theme(panel.grid = element_blank()) +
facet_wrap(~ name, scales = "free")
```

Here’s how to compute the hazard ratio for a 2-unit difference in `pas`

.

`exp(fixef(fit11.9)[7, 1] * 2)`

`## [1] 2.436097`

Here’s what that looks like in a plot.

```
as_draws_df(fit11.9) %>%
transmute(`log-odds` = b_pas,
`hazard ratio` = exp(b_pas),
`hr for a 2-unit difference` = exp(b_pas * 2)) %>%
pivot_longer(everything()) %>%
mutate(name = factor(name,
levels = c("log-odds", "hazard ratio", "hr for a 2-unit difference"))) %>%
ggplot(aes(x = value, y = 0)) +
stat_halfeye(.width = c(.5, .95), normalize = "panels") +
scale_y_continuous(NULL, breaks = NULL) +
xlab("marginal posterior for pas") +
theme(panel.grid = element_blank()) +
facet_wrap(~ name, scales = "free")
```

### 11.4.4 Polytomous substantive predictors.

Unfortunately, neither the `sex`

nor the `sex_pp`

data sets contain the polytomous version of `pt`

as Singer described in this section. However, we can simulate a similar set of dummy variables which will allow us to trace the basic steps in Singer and Willett’s workflow.

Since we’ve been working with the `sex_pp`

data for the past few models, we’ll continue using it here. However, this creates a minor challenge. What we want to do is use the `sample()`

function to randomly assign cases with values of 1, 2, or 3 conditional on whether `pt == 0`

. The catch is, we need to make sure that random value is constant for each case. Our solution will be to first nest the data such that each case only has one row. Here’s what that looks like.

```
set.seed(11)
<-
sex_pp %>%
sex_pp nest(data = c(period, event, d7, d8, d9, d10, d11, d12, pas)) %>%
mutate(random = sample(1:3, size = n(), replace = T)) %>%
mutate(pt_cat = ifelse(pt == 0, pt, random)) %>%
mutate(pt1 = ifelse(pt_cat == 1, 1, 0),
pt2 = ifelse(pt_cat == 2, 1, 0),
pt3 = ifelse(pt_cat == 3, 1, 0)) %>%
select(id, pt, random, pt_cat:pt3, data)
%>%
sex_pp head()
```

```
## # A tibble: 6 × 8
## id pt random pt_cat pt1 pt2 pt3 data
## <dbl> <dbl> <int> <dbl> <dbl> <dbl> <dbl> <list>
## 1 1 0 2 0 0 0 0 <tibble [3 × 9]>
## 2 2 1 2 2 0 1 0 <tibble [6 × 9]>
## 3 3 0 1 0 0 0 0 <tibble [6 × 9]>
## 4 5 1 1 1 1 0 0 <tibble [6 × 9]>
## 5 6 0 1 0 0 0 0 <tibble [5 × 9]>
## 6 7 1 2 2 0 1 0 <tibble [3 × 9]>
```

Here are the number of cases for each of the four levels of our pseudovariable `pt_cat`

.

```
%>%
sex_pp count(pt_cat)
```

```
## # A tibble: 4 × 2
## pt_cat n
## <dbl> <int>
## 1 0 72
## 2 1 41
## 3 2 30
## 4 3 37
```

Our breakdown isn’t exactly the same as the one in the text (p. 391), but we’re in the ballpark. Before we’re ready to fit our next model, we’ll need to `unnest()`

the data, which will transform `sex_pp`

back into the familiar long format.

```
<-
sex_pp %>%
sex_pp unnest(data)
```

Before we fit the model with the `pt*`

dummies, let’s backup and talk about priors, again. When we fit the last four models, our discussion about priors for \(p\) focused on the posterior implications for those parameters in the log-odds space. Things get odd when we consider the implications in the probability space. To demonstrate, let’s simulate from \(\operatorname{Normal}(0, 4)\) and see what it looks like when we transform the draws back into the probability metric.

```
set.seed(11)
tibble(log_odds = rnorm(1e6, mean = 0, sd = 4)) %>%
mutate(p = inv_logit_scaled(log_odds)) %>%
ggplot(aes(x = p)) +
geom_histogram(bins = 50) +
scale_y_continuous(NULL, breaks = NULL) +
theme(panel.grid = element_blank())
```

Perhaps unexpectedly, our log-odds \(\operatorname{Normal}(0, 4)\) prior ended up bunching up the prior mass at the boundaries. Depending on what we want, this may or may not make sense. If we want to regularize the coefficients toward zero in the probability space, something closer to \(\operatorname{Normal}(0, 1)\) would be a better idea. Here’s a look at what happens when we compare three simulations in that range.

```
set.seed(11)
tibble(sd = c(2, 1.5, 1)) %>%
mutate(log_odds = map(sd, ~rnorm(1e6, mean = 0, sd = .))) %>%
unnest(log_odds) %>%
mutate(sd = str_c("sd = ", sd),
p = inv_logit_scaled(log_odds)) %>%
ggplot(aes(x = p)) +
geom_histogram(bins = 50) +
scale_y_continuous(NULL, breaks = NULL) +
theme(panel.grid = element_blank()) +
facet_wrap(~ sd)
```

It looks like the log-odds \(\operatorname{Normal}(0, 1)\) gently regularizes \(p\) toward .5, but still allows for stronger values. This might be a good prior to use for our substantive covariates, what Singer and Willett referred to as the \(\beta\) parameters. But I don’t know that this makes sense for our \(\alpha\) parameters. If you’ve been following along with the model output, the life tables, and so on, you’ll have noticed those tend to drift toward the lower end of the probability range. Regularizing toward \(p = .5\) might not be a good idea. In the absence of good substantive or statistical theory, perhaps it’s best to use a flat prior. The log-odds \(\operatorname{Normal} (0, 1.5)\) prior is nearly flat on the probability space, but it does still push the mass away from the boundaries.

Maybe we can come up with something better. What if we simulated a large number of draws from the \(\operatorname{Uniform}(0, 1)\) distribution, converted those draws to the log-odds metric, and fit a simple Student’s \(t\) model? If we wanted to stay within the Student-\(t\) family of priors, of which the normal is a special case, that would give us a sense of the what prior values would approximate a uniform distribution on the probability scale.

```
set.seed(11)
<-
dat tibble(p = runif(1e5, 0, 1)) %>%
mutate(g = log_odds(p))
.11 <-
fit11brm(data = dat,
family = student,
~ 1,
g chains = 4, cores = 4,
file = "fits/fit11.11")
```

Check the summary.

`print(fit11.11)`

```
## Family: student
## Links: mu = identity; sigma = identity; nu = identity
## Formula: g ~ 1
## Data: dat (Number of observations: 100000)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## Intercept 0.00 0.01 -0.01 0.02 1.00 2592 2306
##
## Family Specific Parameters:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## sigma 1.57 0.01 1.56 1.58 1.00 2155 2743
## nu 7.61 0.17 7.28 7.94 1.00 1923 2551
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

Now we can reverse the process. Here’s what it would look like if we simulated from the Student \(t\)-distribution based on those posterior means and then converted the results into the probability metric.

```
set.seed(11)
tibble(g = rt(1e6, df = 7.61) * 1.57) %>%
mutate(p = inv_logit_scaled(g)) %>%
ggplot(aes(x = p)) +
geom_histogram(bins = 50) +
scale_y_continuous(NULL, breaks = NULL) +
theme(panel.grid = element_blank())
```

Now here’s what it might look like if fit the next model with the \(\operatorname{Normal}(0, 1)\) prior for the \(\beta\) parameters and \(\text{Student-} t (7.61, 0, 1.57)\) prior for the \(\alpha\) parameters.

```
.12 <-
fit11brm(data = sex_pp,
family = binomial,
| trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt1 + pt2 + pt3,
event prior = c(prior(student_t(7.61, 0, 1.57), class = b),
prior(normal(0, 1), class = b, coef = pt1),
prior(normal(0, 1), class = b, coef = pt2),
prior(normal(0, 1), class = b, coef = pt3)),
chains = 4, cores = 4, iter = 2000, warmup = 1000,
seed = 11,
file = "fits/fit11.12")
```

Here is the model summary.

`print(fit11.12)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt1 + pt2 + pt3
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -2.84 0.30 -3.47 -2.27 1.00 4056 2801
## d8 -3.49 0.40 -4.34 -2.79 1.00 5087 2533
## d9 -2.13 0.26 -2.65 -1.65 1.00 3731 2963
## d10 -1.66 0.24 -2.15 -1.20 1.00 3672 2985
## d11 -1.50 0.25 -2.01 -1.01 1.00 4069 2804
## d12 -1.04 0.26 -1.56 -0.54 1.00 4511 3143
## pt1 0.47 0.26 -0.03 0.97 1.00 3423 3383
## pt2 0.52 0.27 -0.01 1.04 1.00 3571 3392
## pt3 1.01 0.26 0.52 1.53 1.00 3168 2937
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

We can use the `tidy()`

function and a few lines of wrangling code to make a version of the table in the middle of page 391.

```
fixef(fit11.12)[7:9, ] %>%
data.frame() %>%
rownames_to_column("predictor") %>%
mutate(`odds ratio` = exp(Estimate)) %>%
select(predictor, Estimate, `odds ratio`) %>%
mutate_if(is.double, round, digits = 3)
```

```
## predictor Estimate odds ratio
## 1 pt1 0.474 1.606
## 2 pt2 0.522 1.685
## 3 pt3 1.013 2.754
```

Because our data did not include the original values for `pt1`

through `pt3`

, the results in our table will not match those in the text. We did get pretty close, though, eh? Hopefully this gives a sense of the workflow.

## 11.5 Displaying fitted hazard and survivor functions

This will be an extension of what we’ve already been doing.

### 11.5.1 A strategy for a single categorical substantive predictor.

We can make our version of Table 11.5 like so. To reduce clutter, we will use abbreviated column names.

```
tibble(time = 7:12,
alpha = fixef(fit11.8)[1:6, 1],
beta = fixef(fit11.8)[7, 1]) %>%
mutate(lh0 = alpha,
lh1 = alpha + beta) %>%
mutate(h0 = inv_logit_scaled(lh0),
h1 = inv_logit_scaled(lh1)) %>%
mutate(s0 = cumprod(1 - h0),
s1 = cumprod(1 - h1)) %>%
# this just simplifies the output
mutate_if(is.double, round, digits = 4)
```

```
## # A tibble: 6 × 9
## time alpha beta lh0 lh1 h0 h1 s0 s1
## <int> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl>
## 1 7 -3.02 0.863 -3.02 -2.15 0.0467 0.104 0.953 0.896
## 2 8 -3.73 0.863 -3.73 -2.87 0.0234 0.0537 0.931 0.848
## 3 9 -2.29 0.863 -2.29 -1.43 0.0916 0.193 0.846 0.684
## 4 10 -1.83 0.863 -1.83 -0.968 0.138 0.275 0.729 0.496
## 5 11 -1.65 0.863 -1.65 -0.791 0.160 0.312 0.612 0.341
## 6 12 -1.18 0.863 -1.18 -0.320 0.234 0.421 0.468 0.198
```

For the `alpha`

and `beta`

columns, we just subset the values from `fixef()`

. The two logit-hazard columns, `lh0`

and `lh1`

, were simple algebraic transformations of `alpha`

and `beta`

, respectively. To make the two hazard columns, `h0`

and `h1`

, we applied the `inv_logit_scaled()`

function to `lh0`

and `lh1`

, respectively. To make the two survival columns, `s0`

and `s1`

, we applied the `cumprod()`

function to one minus the two hazard columns. Note how all this is based off of the posterior means. There’s enough going on with Table 11.5 that it makes sense to ignore uncertainty But when we’re ready to go beyond table glancing and actually make a plot, we will go beyond posterior means and reintroduce the uncertainty in the model.

Two of these plots are quite similar to two of the subplots from Figure 11.4, back in Section 11.2.1. But recall that though those plots were based on `fit11.6`

, which was based on the aggregated data, the plots we are about to make will be based on `fit11.8`

, the analogous model based on the disaggregated person-period data. Regardless of whether the logistic regression model is based on aggregated data, the post-processing approach will involve the `fitted()`

function. However, the specifics of how we use `fitted()`

will differ. For the disaggregated data used to fit `fit11.8`

, here is how we might define the `newdata`

, pump it through the model via `fitted()`

, and wrangle.

```
<-
nd crossing(pt = 0:1,
period = 7:12) %>%
mutate(d7 = if_else(period == 7, 1, 0),
d8 = if_else(period == 8, 1, 0),
d9 = if_else(period == 9, 1, 0),
d10 = if_else(period == 10, 1, 0),
d11 = if_else(period == 11, 1, 0),
d12 = if_else(period == 12, 1, 0))
<-
f fitted(fit11.8,
newdata = nd,
scale = "linear") %>%
data.frame() %>%
bind_cols(nd) %>%
mutate(pt = str_c("pt = ", pt))
f
```

```
## Estimate Est.Error Q2.5 Q97.5 pt period d7 d8 d9 d10 d11 d12
## 1 -3.0158100 0.3226729 -3.6548961 -2.4101270 pt = 0 7 1 0 0 0 0 0
## 2 -3.7332639 0.4320540 -4.6605608 -2.9515943 pt = 0 8 0 1 0 0 0 0
## 3 -2.2948031 0.2781025 -2.8705837 -1.7807053 pt = 0 9 0 0 1 0 0 0
## 4 -1.8307105 0.2561879 -2.3418476 -1.3360186 pt = 0 10 0 0 0 1 0 0
## 5 -1.6544779 0.2747193 -2.2144784 -1.1401863 pt = 0 11 0 0 0 0 1 0
## 6 -1.1834418 0.2682602 -1.7184945 -0.6706097 pt = 0 12 0 0 0 0 0 1
## 7 -2.1526553 0.2819101 -2.7254006 -1.6252754 pt = 1 7 1 0 0 0 0 0
## 8 -2.8701093 0.4043764 -3.7196783 -2.1488898 pt = 1 8 0 1 0 0 0 0
## 9 -1.4316484 0.2357858 -1.9315490 -0.9821854 pt = 1 9 0 0 1 0 0 0
## 10 -0.9675558 0.2256333 -1.4117138 -0.5270430 pt = 1 10 0 0 0 1 0 0
## 11 -0.7913233 0.2466043 -1.2877582 -0.3239337 pt = 1 11 0 0 0 0 1 0
## 12 -0.3202872 0.2561600 -0.8262651 0.1804904 pt = 1 12 0 0 0 0 0 1
```

Here we make and save the upper two panels of Figure 11.6.

```
# logit(hazard)
<-
p1 %>%
f
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "fitted logit(hazard)",
y = NULL) +
coord_cartesian(ylim = c(-4, 0)) +
theme(legend.background = element_rect(fill = "transparent"),
legend.key = element_rect(color = "grey92"),
legend.position = c(.1, .825))
# hazard
<-
p2 %>%
f mutate_at(vars(Estimate, Q2.5, Q97.5), .funs = inv_logit_scaled) %>%
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "fitted hazard",
y = NULL) +
coord_cartesian(ylim = c(0, .5)) +
theme(legend.position = "none")
```

Before we’re ready to make the last panel, we’ll redo our `fitted()`

work, this time including predicted values for grade 6.

```
<-
nd crossing(pt = 0:1,
period = 6:12) %>%
mutate(d6 = if_else(period == 6, 1, 0),
d7 = if_else(period == 7, 1, 0),
d8 = if_else(period == 8, 1, 0),
d9 = if_else(period == 9, 1, 0),
d10 = if_else(period == 10, 1, 0),
d11 = if_else(period == 11, 1, 0),
d12 = if_else(period == 12, 1, 0))
<-
f fitted(fit11.8,
newdata = nd) %>%
data.frame() %>%
bind_cols(nd) %>%
mutate(pt = str_c("pt = ", pt))
f
```

```
## Estimate Est.Error Q2.5 Q97.5 pt period d6 d7 d8 d9 d10 d11 d12
## 1 0.50000000 0.00000000 0.500000000 0.50000000 pt = 0 6 1 0 0 0 0 0 0
## 2 0.04882018 0.01493324 0.025212098 0.08240372 pt = 0 7 0 1 0 0 0 0 0
## 3 0.02536701 0.01037586 0.009372481 0.04966121 pt = 0 8 0 0 1 0 0 0 0
## 4 0.09415150 0.02331905 0.053627034 0.14421607 pt = 0 9 0 0 0 1 0 0 0
## 5 0.14095429 0.03084918 0.087715959 0.20816556 pt = 0 10 0 0 0 0 1 0 0
## 6 0.16390446 0.03708717 0.098457842 0.24228622 pt = 0 11 0 0 0 0 0 1 0
## 7 0.23779436 0.04790263 0.152065178 0.33836034 pt = 0 12 0 0 0 0 0 0 1
## 8 0.70132350 0.04546198 0.606327819 0.78490363 pt = 1 6 1 0 0 0 0 0 0
## 9 0.10699011 0.02662588 0.061491062 0.16447861 pt = 1 7 0 1 0 0 0 0 0
## 10 0.05728882 0.02112107 0.023668020 0.10443502 pt = 1 8 0 0 1 0 0 0 0
## 11 0.19546013 0.03629156 0.126579224 0.27245837 pt = 1 9 0 0 0 1 0 0 0
## 12 0.27761544 0.04480749 0.195963894 0.37120682 pt = 1 10 0 0 0 0 1 0 0
## 13 0.31430028 0.05231257 0.216232504 0.41971737 pt = 1 11 0 0 0 0 0 1 0
## 14 0.42184774 0.06152185 0.304435381 0.54500051 pt = 1 12 0 0 0 0 0 0 1
```

The values for grade 6 (i.e., those for when `d6 == 1`

) are nonsensical. The main reason we included `d6`

in the fitted results and in the `nd`

data is so we’d have the slots in our `f`

object. In the code block below, we’ll fill those slots with the appropriate values (`0`

) and then convert the hazard summaries to the survival (i.e., cumulative probability) metric.

```
<-
f %>%
f mutate(Estimate = if_else(period == 6, 0, Estimate),
Q2.5 = if_else(period == 6, 0, Q2.5),
Q97.5 = if_else(period == 6, 0, Q97.5)) %>%
group_by(pt) %>%
mutate(s = cumprod(1 - Estimate),
s_lower = cumprod(1 - Q2.5),
s_upper = cumprod(1 - Q97.5)) %>%
select(pt:d12, s:s_upper)
%>% glimpse() f
```

```
## Rows: 14
## Columns: 12
## Groups: pt [2]
## $ pt <chr> "pt = 0", "pt = 0", "pt = 0", "pt = 0", "pt = 0", "pt = 0", "pt = 0", "pt = 1", "pt = 1", "p…
## $ period <int> 6, 7, 8, 9, 10, 11, 12, 6, 7, 8, 9, 10, 11, 12
## $ d6 <dbl> 1, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0
## $ d7 <dbl> 0, 1, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0
## $ d8 <dbl> 0, 0, 1, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0
## $ d9 <dbl> 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 1, 0, 0, 0
## $ d10 <dbl> 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 1, 0, 0
## $ d11 <dbl> 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 1, 0
## $ d12 <dbl> 0, 0, 0, 0, 0, 0, 1, 0, 0, 0, 0, 0, 0, 1
## $ s <dbl> 1.0000000, 0.9511798, 0.9270512, 0.8397680, 0.7213991, 0.6031585, 0.4597308, 1.0000000, 0.89…
## $ s_lower <dbl> 1.0000000, 0.9747879, 0.9656517, 0.9138667, 0.8337060, 0.7516211, 0.6373257, 1.0000000, 0.93…
## $ s_upper <dbl> 1.00000000, 0.91759628, 0.87202734, 0.74626698, 0.59091990, 0.44774815, 0.29624793, 1.000000…
```

Make and save the final panel.

```
# save the interpolated median lifetime values
<- c(make_iml(lt %>% filter(pt == "pt = 0")), make_iml(lt %>% filter(pt == "pt = 1")))
imls
# hazard
<-
p3 %>%
f
ggplot(aes(x = period, group = pt,
fill = pt, color = pt)) +
geom_hline(yintercept = .5, color = "white") +
geom_segment(x = imls[1], xend = imls[1],
y = -Inf, yend = .5,
color = "white", linetype = 2) +
geom_segment(x = imls[2], xend = imls[2],
y = -Inf, yend = .5,
color = "white", linetype = 2) +
geom_ribbon(aes(ymin = s_lower, ymax = s_upper),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = s)) +
scale_y_continuous(NULL, breaks = c(0, .5, 1)) +
labs(subtitle = "fitted survival probability") +
coord_cartesian(ylim = c(0, 1)) +
theme(legend.position = "none")
```

Combine the subplots to finish off our version of Figure 11.6.

```
/ p2 / p3) &
(p1 scale_fill_viridis_d(NULL, option = "A", end = .6) &
scale_color_viridis_d(NULL, option = "A", end = .6) &
scale_x_continuous("Grade", breaks = 6:12, limits = c(6, 12)) &
theme(panel.grid = element_blank())
```

Here is the breakdown of what percentage of boys will still be virgins at grades 9 and 12, based on `pt`

status, as indicated by `fit11.8`

.

```
%>%
f filter(period %in% c(9, 12)) %>%
mutate_if(is.double, ~ (. * 100) %>% round(digits = 0)) %>%
mutate(`percent virgins` = str_c(s, " [", s_lower, ", ", s_upper, "]")) %>%
select(period, pt, `percent virgins`) %>%
arrange(period)
```

```
## # A tibble: 4 × 3
## # Groups: pt [2]
## period pt `percent virgins`
## <int> <chr> <chr>
## 1 9 pt = 0 84 [91, 75]
## 2 9 pt = 1 68 [80, 54]
## 3 12 pt = 0 46 [64, 30]
## 4 12 pt = 1 19 [35, 9]
```

### 11.5.2 Extending this strategy to multiple predictors (some of which are continuous).

It is easy to display fitted hazard and survivor functions for model involving multiple predictor by extending these ideas in a straightforward manner. Instead of plotting one fitted function for

eachpredictor value, select severalprototypicalpredictor values (using strategies presented in section 4.5.3 and plot fitted functions for combinations of these values. (p. 394,emphasisin the original)

We’ll be focusing on `fit11.10`

, which includes both `pt`

and `sas`

as substantive predictors. `pt`

only takes two values, 0 and 1. For pas, we’ll use the conventional -1, 0, and 1. Here’s the `fitted()`

-related code.

```
<-
nd crossing(pt = 0:1,
pas = -1:1) %>%
expand(nesting(pt, pas),
period = 6:12) %>%
mutate(d6 = if_else(period == 6, 1, 0),
d7 = if_else(period == 7, 1, 0),
d8 = if_else(period == 8, 1, 0),
d9 = if_else(period == 9, 1, 0),
d10 = if_else(period == 10, 1, 0),
d11 = if_else(period == 11, 1, 0),
d12 = if_else(period == 12, 1, 0))
<-
f fitted(fit11.10,
newdata = nd) %>%
data.frame() %>%
bind_cols(nd)
head(f)
```

```
## Estimate Est.Error Q2.5 Q97.5 pt pas period d6 d7 d8 d9 d10 d11 d12
## 1 0.42502258 0.030246749 0.367418550 0.48511371 0 -1 6 1 0 0 0 0 0 0
## 2 0.04077901 0.013144117 0.019410454 0.07067494 0 -1 7 0 1 0 0 0 0 0
## 3 0.02119052 0.008558962 0.008440147 0.04130416 0 -1 8 0 0 1 0 0 0 0
## 4 0.08141420 0.020968559 0.046732027 0.12872256 0 -1 9 0 0 0 1 0 0 0
## 5 0.12326596 0.029408292 0.072767576 0.18858903 0 -1 10 0 0 0 0 1 0 0
## 6 0.14293494 0.034472676 0.082039623 0.21718230 0 -1 11 0 0 0 0 0 1 0
```

Make the two subplots.

```
# logit(hazard)
<-
p1 %>%
f mutate(pt = str_c("pt = ", pt),
pas = str_c("pas = ", pas)) %>%
mutate(pas = factor(pas,
levels = str_c("pas = ", 1:-1))) %>%
filter(period > 6) %>%
ggplot(aes(x = period, group = pas,
fill = pas, color = pas)) +
geom_ribbon(aes(ymin = Q2.5, ymax = Q97.5),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = Estimate)) +
labs(subtitle = "fitted logit(hazard)",
y = NULL) +
coord_cartesian(ylim = c(0, .5)) +
facet_wrap(~ pt)
# hazard
<-
p2 %>%
f mutate(Estimate = if_else(period == 6, 0, Estimate),
Q2.5 = if_else(period == 6, 0, Q2.5),
Q97.5 = if_else(period == 6, 0, Q97.5)) %>%
mutate(pt = str_c("pt = ", pt),
pas = str_c("pas = ", pas)) %>%
mutate(pas = factor(pas,
levels = str_c("pas = ", 1:-1))) %>%
group_by(pt, pas) %>%
mutate(s = cumprod(1 - Estimate),
s_lower = cumprod(1 - Q2.5),
s_upper = cumprod(1 - Q97.5)) %>%
ggplot(aes(x = period, group = pas,
fill = pas, color = pas)) +
geom_hline(yintercept = .5, color = "white") +
geom_ribbon(aes(ymin = s_lower, ymax = s_upper),
linewidth = 0, alpha = 1/6) +
geom_line(aes(y = s)) +
scale_y_continuous(NULL, breaks = c(0, .5, 1)) +
labs(subtitle = "fitted survival probability") +
coord_cartesian(ylim = c(0, 1)) +
theme(legend.position = "none") +
facet_wrap(~ pt)
```

Combine the subplots to make our version of Figure 11.7.

```
/ p2) &
((p1 scale_fill_viridis_d(NULL, option = "D", end = .8, direction = -1) &
scale_color_viridis_d(NULL, option = "D", end = .8, direction = -1) &
scale_x_continuous("Grade", breaks = 6:12, limits = c(6, 12)) &
theme(panel.grid = element_blank())) +
plot_layout(guides = "collect")
```

Here we departed from the text a bit by separating the subplots by `pt`

status. They’re already cluttered enough as is.

### 11.5.3 Two cautions when interpreting fitted hazard and survivor functions.

Beware of inferring statistical interaction of a substantive predictor and time when examining plots if fitted hazard and survivor functions. The root of this difficulty is in our use of a link function.

Because the model expresses the linear effect of the predictor on logit hazard, you cannot draw a conclusion about the stability of an effect using graphs plotted on a raw hazard scale. In fact, the logic works in the

oppositedirection. If the size of the gap between fitted hazard functions is constant over time, [the] effect of the predictor mustvaryover time! (pp. 396-397,emphasisin the original)

Also, please don’t confuse plots of fitted values with descriptive sample-based plots. Hopefully our inclusion of 95% intervals helps prevent this.

## 11.6 Comparing models using ~~deviance statistics and~~ information criteria

We now introduce two important questions that we usually address before interpreting parameters and displaying results: Which of the alternative models fits better: Might a predictor’s observed effect be the result of nothing more than sampling variation? (p. 397)

Much of the material in this section will be a refresher from the material we covered in Section 4.6.

### 11.6.1 The deviance statistic.

The log-likelihood, LL, is

a summary statistic routinely output (in some form) by any program that provides ML estimates. As discussed in section 4.6, its relative magnitude across a series of models fit to the same set of data can be informative (although its absolute magnitude is not). The larger the LL statistic, the better the fit. (pp. 397–398)

Note that *in some form* part. Frequentist software typically returns the LL for a given model as a single value. As we learned way back in Section 4.6, we can use the `log_lik()`

function to get the LL information from our **brms** fits. However, form the **brms** reference manual we discover `log_lik()`

returns an “S x N matrix containing the pointwise log-likelihood samples, where S is the number of samples and N is the number of observations in the data” (p. 112). Using `fit11.7`

as a test case, here’s what that looks like.

```
log_lik(fit11.7) %>%
str()
```

```
## num [1:4000, 1:822] -0.0988 -0.0798 -0.088 -0.0665 -0.0827 ...
## - attr(*, "dimnames")=List of 2
## ..$ : NULL
## ..$ : NULL
```

To compute the LL for each HMC iteration, you sum across the rows. Deviance is that sum multiplied by -2. Here’s that in a tibble.

```
<-
ll .7 %>%
fit11log_lik() %>%
as_tibble(.name_repair = ~ str_c("c", 1:822)) %>%
mutate(ll = rowSums(.)) %>%
mutate(deviance = -2 * ll) %>%
select(ll, deviance, everything())
ll
```

```
## # A tibble: 4,000 × 824
## ll deviance c1 c2 c3 c4 c5 c6 c7 c8 c9 c10 c11 c12
## <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl> <dbl>
## 1 -330. 660. -0.0988 -0.0477 -2.03 -0.0988 -0.0477 -0.141 -0.252 -0.225 -0.234 -0.0988 -0.0477 -0.141
## 2 -328. 655. -0.0798 -0.0435 -1.91 -0.0798 -0.0435 -0.160 -0.201 -0.348 -0.462 -0.0798 -0.0435 -0.160
## 3 -331. 661. -0.0880 -0.0230 -1.86 -0.0880 -0.0230 -0.170 -0.210 -0.298 -0.620 -0.0880 -0.0230 -0.170
## 4 -333. 667. -0.0665 -0.0525 -2.35 -0.0665 -0.0525 -0.0998 -0.182 -0.343 -0.588 -0.0665 -0.0525 -0.0998
## 5 -337. 675. -0.0827 -0.0564 -2.33 -0.0827 -0.0564 -0.102 -0.140 -0.440 -0.556 -0.0827 -0.0564 -0.102
## 6 -328. 657. -0.0779 -0.0514 -2.24 -0.0779 -0.0514 -0.113 -0.205 -0.237 -0.376 -0.0779 -0.0514 -0.113
## 7 -329. 658. -0.0647 -0.0597 -2.03 -0.0647 -0.0597 -0.140 -0.201 -0.358 -0.359 -0.0647 -0.0597 -0.140
## 8 -327. 654. -0.0931 -0.0428 -1.72 -0.0931 -0.0428 -0.198 -0.241 -0.317 -0.349 -0.0931 -0.0428 -0.198
## 9 -327. 653. -0.0934 -0.0483 -2.04 -0.0934 -0.0483 -0.139 -0.254 -0.278 -0.443 -0.0934 -0.0483 -0.139
## 10 -329. 658. -0.0837 -0.0331 -2.19 -0.0837 -0.0331 -0.119 -0.191 -0.332 -0.363 -0.0837 -0.0331 -0.119
## # ℹ 3,990 more rows
## # ℹ 810 more variables: c13 <dbl>, c14 <dbl>, c15 <dbl>, c16 <dbl>, c17 <dbl>, c18 <dbl>, c19 <dbl>,
## # c20 <dbl>, c21 <dbl>, c22 <dbl>, c23 <dbl>, c24 <dbl>, c25 <dbl>, c26 <dbl>, c27 <dbl>, c28 <dbl>,
## # c29 <dbl>, c30 <dbl>, c31 <dbl>, c32 <dbl>, c33 <dbl>, c34 <dbl>, c35 <dbl>, c36 <dbl>, c37 <dbl>,
## # c38 <dbl>, c39 <dbl>, c40 <dbl>, c41 <dbl>, c42 <dbl>, c43 <dbl>, c44 <dbl>, c45 <dbl>, c46 <dbl>,
## # c47 <dbl>, c48 <dbl>, c49 <dbl>, c50 <dbl>, c51 <dbl>, c52 <dbl>, c53 <dbl>, c54 <dbl>, c55 <dbl>,
## # c56 <dbl>, c57 <dbl>, c58 <dbl>, c59 <dbl>, c60 <dbl>, c61 <dbl>, c62 <dbl>, c63 <dbl>, c64 <dbl>, …
```

Since we have distributions for the LL and deviance, we may as well visualize them in a plot.

```
%>%
ll pivot_longer(ll:deviance) %>%
mutate(name = factor(name, levels = c("ll", "deviance"))) %>%
ggplot(aes(x = value, y = 0)) +
stat_halfeye(point_interval = median_qi, .width = .95, normalize = "panels") +
scale_y_continuous(NULL, breaks = NULL) +
xlab(NULL) +
theme(panel.grid = element_blank()) +
facet_wrap(~ name, scales = "free")
```

Here’s how to compute the LL and deviance distributions for each of our four models, `fit11.7`

through `fit11.10`

, in bulk.

```
<-
ll tibble(model = str_c("model ", letters[1:4]),
name = str_c("fit11.", 7:10)) %>%
mutate(fit = map(name, get)) %>%
mutate(ll = map(fit, ~log_lik(.) %>% data.frame() %>% transmute(ll = rowSums(.)))) %>%
select(-fit) %>%
unnest(ll) %>%
mutate(deviance = -2 * ll)
%>%
ll glimpse()
```

```
## Rows: 16,000
## Columns: 4
## $ model <chr> "model a", "model a", "model a", "model a", "model a", "model a", "model a", "model a", "mo…
## $ name <chr> "fit11.7", "fit11.7", "fit11.7", "fit11.7", "fit11.7", "fit11.7", "fit11.7", "fit11.7", "fi…
## $ ll <dbl> -329.7851, -327.7458, -330.6678, -333.3882, -337.4965, -328.3698, -328.9242, -326.9331, -32…
## $ deviance <dbl> 659.5702, 655.4915, 661.3355, 666.7763, 674.9931, 656.7396, 657.8484, 653.8662, 653.2041, 6…
```

Now plot the LL and deviance distributions for each.

```
%>%
ll pivot_longer(ll:deviance,
names_to = "statistic") %>%
mutate(statistic = factor(statistic, levels = c("ll", "deviance"))) %>%
ggplot(aes(x = value, y = model)) +
stat_halfeye(point_interval = median_qi, .width = .95, normalize = "panels") +
labs(x = NULL,
y = NULL) +
theme(panel.grid = element_blank()) +
facet_wrap(~ statistic, scales = "free_x")
```

### 11.6.2 Deviance-based hypothesis tests for individual predictors.

Singer and Willett wrote: “Comparing deviance statistics for pairs of nested models that differ only by a single substantive predictor permits evaluation of the ‘statistical significance’ of that predictor” (p. 399). I’m just not going to appeal to null-hypothesis significance testing in this project and, as an extension, I am not going to appeal to tests using the \(\chi^2\) distribution. But sure, you could take our deviance distributions and compare them with difference distributions. Singer and Willett made four deviance comparisons in this section. Here’s what that might look like using our deviance distributions.

```
%>%
ll select(model, deviance) %>%
mutate(iter = rep(1:4000, times = 4)) %>%
pivot_wider(names_from = model,
values_from = deviance) %>%
mutate(`a - b` = `model a` - `model b`,
`a - c` = `model a` - `model c`,
`c - d` = `model c` - `model d`,
`b - d` = `model b` - `model d`) %>%
pivot_longer(contains("-")) %>%
ggplot(aes(x = value, y = name)) +
stat_halfeye(point_interval = median_qi, .width = .95, normalize = "panels") +
labs(x = "deviance difference distribution",
y = NULL) +
theme(panel.grid = element_blank())
```

But really, like no one does this with Bayesian models. If you think you have a good theoretical reason to use this approach, do not cite this project as a justification. I do not endorse it.

### 11.6.4 Comparing nonnested models using [WAIC and LOO].

Now we return to our preferred methods for model comparison. Use the `add_criterion()`

function to compute the WAIC and LOO and add their output to the model fits.

```
.7 <- add_criterion(fit11.7, c("loo", "waic"))
fit11.8 <- add_criterion(fit11.8, c("loo", "waic"))
fit11.9 <- add_criterion(fit11.9, c("loo", "waic"))
fit11.10 <- add_criterion(fit11.10, c("loo", "waic")) fit11
```

First compare Models B and C (i.e., `fit11.8`

and `fit11.9`

, respectively).

`loo_compare(fit11.8, fit11.9, criterion = "loo") %>% print(simplify = F)`

```
## elpd_diff se_diff elpd_loo se_elpd_loo p_loo se_p_loo looic se_looic
## fit11.8 0.0 0.0 -324.5 17.6 7.2 0.6 649.0 35.3
## fit11.9 -1.1 4.6 -325.6 17.6 7.0 0.6 651.3 35.1
```

`loo_compare(fit11.8, fit11.9, criterion = "waic") %>% print(simplify = F)`

```
## elpd_diff se_diff elpd_waic se_elpd_waic p_waic se_p_waic waic se_waic
## fit11.8 0.0 0.0 -324.5 17.6 7.2 0.6 649.0 35.3
## fit11.9 -1.1 4.6 -325.6 17.6 7.0 0.6 651.2 35.1
```

In a head-to-head comparison, Model B is a little better than Model C. However, the standard error for their difference score is about three times as large as the difference itself. This is not a difference I would write home about. Now compare Models A through D.

`loo_compare(fit11.7, fit11.8, fit11.9, fit11.10, criterion = "loo") %>% print(simplify = F)`

```
## elpd_diff se_diff elpd_loo se_elpd_loo p_loo se_p_loo looic se_looic
## fit11.10 0.0 0.0 -322.8 17.6 8.3 0.6 645.6 35.1
## fit11.8 -1.7 2.6 -324.5 17.6 7.2 0.6 649.0 35.3
## fit11.9 -2.8 2.8 -325.6 17.6 7.0 0.6 651.3 35.1
## fit11.7 -9.4 4.8 -332.2 17.6 6.2 0.5 664.3 35.3
```

`loo_compare(fit11.7, fit11.8, fit11.9, fit11.10, criterion = "waic") %>% print(simplify = F)`

```
## elpd_diff se_diff elpd_waic se_elpd_waic p_waic se_p_waic waic se_waic
## fit11.10 0.0 0.0 -322.8 17.6 8.2 0.6 645.6 35.1
## fit11.8 -1.7 2.6 -324.5 17.6 7.2 0.6 649.0 35.3
## fit11.9 -2.8 2.8 -325.6 17.6 7.0 0.6 651.2 35.1
## fit11.7 -9.4 4.8 -332.2 17.6 6.2 0.5 664.3 35.3
```

Model D (i.e., `fit11.10`

, the full model) has the best (i.e., lowest) WAIC and LOO estimates. However, the standard errors for its difference scores with the other models is on the large side, particularly for Models B and C. So sure, adding either `pt`

or `sas`

to the model helps a bit and adding them both helps a little more, but neither predictor is a huge winner when you take that model complexity penalty into account.

As discussed earlier, we can also compare the models using weights. Here we’ll use the WAIC, LOO, and stacking weights to compare all four models.

`model_weights(fit11.7, fit11.8, fit11.9, fit11.10, weights = "loo") %>% round(digits = 3)`

```
## fit11.7 fit11.8 fit11.9 fit11.10
## 0.000 0.146 0.048 0.805
```

`model_weights(fit11.7, fit11.8, fit11.9, fit11.10, weights = "waic") %>% round(digits = 3)`

```
## fit11.7 fit11.8 fit11.9 fit11.10
## 0.000 0.146 0.048 0.806
```

`model_weights(fit11.7, fit11.8, fit11.9, fit11.10, weights = "stacking") %>% round(digits = 3)`

```
## fit11.7 fit11.8 fit11.9 fit11.10
## 0.000 0.382 0.292 0.326
```

Model D has the best showing across the three weighting schemes.

## 11.7 Statistical inference using [uncertainty in the Bayesian posterior]

I generally take a model-based approach to Bayesian statistics and I prefer to scrutinize marginal posteriors, consider effect sizes, and use graphical depictions of my models (e.g., posterior predictive checks) over hypothesis testing. Further extending that approach, here, puts us at further odds with the content in the test. In addition, the authors spent some time discussing the asymptotic properties of ML standard errors. Our Bayesian approach is not based on asymptotic theory and we just don’t need to concern ourselves with whether our marginal posteriors are Gaussian. They often are, which is nice. But we summarize our posteriors with percentile-based 95% intervals, we are not presuming they are symmetric or Gaussian.

### 11.7.1 The ~~Wald chi-square statistic~~.

This will not be our approach. On page 404, Singer and Willett wrote: “The logistic regression analysis routines in all major statistical packages routinely output asymptotic standard errors.” This comment presumes we’re focusing on frequentist packages. Our rough analogue to frequentist standard errors is our Bayesian posterior standard deviations. The authors focused on the two substantive predictors from Model D (i.e., `fit11.10`

). Here’s another look at the **brms** summary.

`print(fit11.10)`

```
## Family: binomial
## Links: mu = logit
## Formula: event | trials(1) ~ 0 + d7 + d8 + d9 + d10 + d11 + d12 + pt + pas
## Data: sex_pp (Number of observations: 822)
## Draws: 4 chains, each with iter = 2000; warmup = 1000; thin = 1;
## total post-warmup draws = 4000
##
## Population-Level Effects:
## Estimate Est.Error l-95% CI u-95% CI Rhat Bulk_ESS Tail_ESS
## d7 -2.91 0.33 -3.58 -2.30 1.00 2602 2358
## d8 -3.61 0.42 -4.47 -2.86 1.00 3035 2921
## d9 -2.15 0.28 -2.72 -1.63 1.00 2148 2448
## d10 -1.69 0.27 -2.23 -1.17 1.00 2187 2907
## d11 -1.52 0.28 -2.11 -0.98 1.00 2639 2409
## d12 -1.01 0.28 -1.58 -0.45 1.00 2571 2681
## pt 0.64 0.24 0.16 1.11 1.01 1358 1532
## pas 0.30 0.12 0.06 0.54 1.00 2849 2810
##
## Draws were sampled using sampling(NUTS). For each parameter, Bulk_ESS
## and Tail_ESS are effective sample size measures, and Rhat is the potential
## scale reduction factor on split chains (at convergence, Rhat = 1).
```

Recall that the second column for our ‘Population-Level Effects’, ‘Est.Error’, contains the standard deviation for each dimension of the posteriors listed (i.e., for each parameter ranging from `d7`

to `pas`

). This is similar, but distinct from, the frequentist standard error. Instead of focusing on \(p\)-values connected to standard errors, why not look at the marginal posteriors directly?

```
<- as_draws_df(fit11.10)
draws
%>%
draws pivot_longer(b_pt:b_pas) %>%
ggplot(aes(x = value, y = name, fill = stat(x > 0))) +
stat_slab() +
scale_fill_manual(values = c("blue3", "red3")) +
labs(x = "marginal posterior",
y = NULL) +
coord_cartesian(ylim = c(1.5, 2)) +
theme(panel.grid = element_blank())
```

If we’d like to keep with the NHST perspective, zero is not a particularly credible value for either parameter. But neither are negative values in generals. In terms of uncertainty, look how much wider the posterior for `pt`

is when compared with `pas`

. And don’t forget that these are on the log-odds scale.

Looking at those densities might lead one to ask, *Exactly what proportion of the posterior draws for each is zero or below*? You can compute that like this.

```
%>%
draws pivot_longer(b_pt:b_pas) %>%
group_by(name) %>%
summarise(`percent zero or below` = 100 * mean(value <= 0))
```

```
## # A tibble: 2 × 2
## name `percent zero or below`
## <chr> <dbl>
## 1 b_pas 0.725
## 2 b_pt 0.375
```

Less that 1% of the draws were zero or below for each.

### 11.7.2 [Asymmetric credible intervals] for parameters and odds ratios.

Whether we use percentile-based credible intervals, as we typically do, or use highest posterior density intervals, neither depends on asymptotic theory nor do they depend on the posterior standard deviation. That is, our Bayesian intervals do not presume the marginal posteriors are Gaussian. Let’s look back at the summary output for `fit11.10`

, this time using the `fixef()`

function.

`fixef(fit11.10)`

```
## Estimate Est.Error Q2.5 Q97.5
## d7 -2.9070130 0.3296742 -3.57689882 -2.3023789
## d8 -3.6113414 0.4193496 -4.47419502 -2.8617981
## d9 -2.1527055 0.2792362 -2.71607751 -1.6276566
## d10 -1.6864101 0.2708240 -2.22509953 -1.1692466
## d11 -1.5164174 0.2829950 -2.10973368 -0.9784899
## d12 -1.0071412 0.2842673 -1.57657489 -0.4529368
## pt 0.6390128 0.2419286 0.15816324 1.1104594
## pas 0.3033367 0.1241903 0.05956277 0.5433073
```

We find the lower- and upper-limits for our percentile-based Bayesian credible intervals in the last two columns. If you’d like HDIs instead, use the convenience functions from **tidybayes**.

```
%>%
draws pivot_longer(b_pt:b_pas) %>%
group_by(name) %>%
mean_hdi(value)
```

```
## # A tibble: 2 × 7
## name value .lower .upper .width .point .interval
## <chr> <dbl> <dbl> <dbl> <dbl> <chr> <chr>
## 1 b_pas 0.303 0.0622 0.544 0.95 mean hdi
## 2 b_pt 0.639 0.148 1.09 0.95 mean hdi
```

We can exponentiate our measures of central tendency (e.g., posterior means) and posterior intervals to transform them out of the log-odds metric an into the odds-ratio metric. Here are the results working directly with `fixef()`

.

`fixef(fit11.10)[c("pt", "pas"), -2] %>% exp()`

```
## Estimate Q2.5 Q97.5
## pt 1.89461 1.171357 3.035753
## pas 1.35437 1.061372 1.721692
```

Keep in mind that fixating on just the 95% intervals is a little NHST-centric. Since we have entire posterior distributions to summarize, we might consider other intervals. Here we use another graphical approach by using `tidybayes_statintervalh()`

to mark off the 10, 30, 50, 70, and 90% intervals for both substantive predictors. Both are in the odds-ratio metric.

```
%>%
draws pivot_longer(b_pt:b_pas) %>%
mutate(`odds ratio` = exp(value)) %>%
ggplot(aes(x = `odds ratio`, y = name)) +
stat_interval(size = 5, .width = seq(from = .1, to = .9, by = .2)) +
scale_color_grey("CI level:", start = .8, end = .2) +
scale_x_continuous(breaks = 1:3) +
ylab(NULL) +
coord_cartesian(xlim = c(1, 3)) +
theme(legend.position = "top",
panel.grid = element_blank())
```

The frequentist 95% confidence intervals are asymmetric when expressed in the odds-ratio metric and so are our various Bayesian intervals. However, the asymmetry in our Bayesian intervals is less noteworthy because there was no explicit assumption of symmetry when they were in the log-odds metric.

## Session info

`sessionInfo()`

```
## R version 4.3.0 (2023-04-21)
## Platform: x86_64-apple-darwin20 (64-bit)
## Running under: macOS Monterey 12.4
##
## Matrix products: default
## BLAS: /Library/Frameworks/R.framework/Versions/4.3-x86_64/Resources/lib/libRblas.0.dylib
## LAPACK: /Library/Frameworks/R.framework/Versions/4.3-x86_64/Resources/lib/libRlapack.dylib; LAPACK version 3.11.0
##
## locale:
## [1] en_US.UTF-8/en_US.UTF-8/en_US.UTF-8/C/en_US.UTF-8/en_US.UTF-8
##
## time zone: America/Chicago
## tzcode source: internal
##
## attached base packages:
## [1] stats graphics grDevices utils datasets methods base
##
## other attached packages:
## [1] tidybayes_3.0.4 brms_2.19.0 Rcpp_1.0.10 patchwork_1.1.2 survival_3.5-5 lubridate_1.9.2
## [7] forcats_1.0.0 stringr_1.5.0 dplyr_1.1.2 purrr_1.0.1 readr_2.1.4 tidyr_1.3.0
## [13] tibble_3.2.1 ggplot2_3.4.2 tidyverse_2.0.0
##
## loaded via a namespace (and not attached):
## [1] svUnit_1.0.6 shinythemes_1.2.0 splines_4.3.0 later_1.3.1
## [5] gamm4_0.2-6 xts_0.13.1 lifecycle_1.0.3 StanHeaders_2.26.25
## [9] processx_3.8.1 lattice_0.21-8 vroom_1.6.3 MASS_7.3-58.4
## [13] crosstalk_1.2.0 ggdist_3.3.0 backports_1.4.1 magrittr_2.0.3
## [17] sass_0.4.6 rmarkdown_2.21 jquerylib_0.1.4 httpuv_1.6.11
## [21] zip_2.3.0 askpass_1.1 pkgbuild_1.4.0 minqa_1.2.5
## [25] multcomp_1.4-23 abind_1.4-5 TH.data_1.1-2 tensorA_0.36.2
## [29] sandwich_3.0-2 gdtools_0.3.3 inline_0.3.19 crul_1.4.0
## [33] xslt_1.4.4 bridgesampling_1.1-2 codetools_0.2-19 DT_0.27
## [37] xml2_1.3.4 tidyselect_1.2.0 bayesplot_1.10.0 httpcode_0.3.0
## [41] farver_2.1.1 lme4_1.1-33 matrixStats_0.63.0 stats4_4.3.0
## [45] base64enc_0.1-3 jsonlite_1.8.4 ellipsis_0.3.2 emmeans_1.8.6
## [49] systemfonts_1.0.4 projpred_2.5.0 tools_4.3.0 ragg_1.2.5
## [53] glue_1.6.2 gridExtra_2.3 xfun_0.39 mgcv_1.8-42
## [57] distributional_0.3.2 loo_2.6.0 withr_2.5.0 fastmap_1.1.1
## [61] boot_1.3-28.1 fansi_1.0.4 shinyjs_2.1.0 openssl_2.0.6
## [65] callr_3.7.3 digest_0.6.31 timechange_0.2.0 R6_2.5.1
## [69] mime_0.12 estimability_1.4.1 textshaping_0.3.6 colorspace_2.1-0
## [73] gtools_3.9.4 markdown_1.7 threejs_0.3.3 utf8_1.2.3
## [77] generics_0.1.3 fontLiberation_0.1.0 data.table_1.14.8 prettyunits_1.1.1
## [81] htmlwidgets_1.6.2 pkgconfig_2.0.3 dygraphs_1.1.1.6 gtable_0.3.3
## [85] htmltools_0.5.5 fontBitstreamVera_0.1.1 bookdown_0.34 scales_1.2.1
## [89] posterior_1.4.1 knitr_1.42 rstudioapi_0.14 tzdb_0.4.0
## [93] reshape2_1.4.4 uuid_1.1-0 coda_0.19-4 checkmate_2.2.0
## [97] nlme_3.1-162 curl_5.0.0 nloptr_2.0.3 cachem_1.0.8
## [101] zoo_1.8-12 flextable_0.9.1 parallel_4.3.0 miniUI_0.1.1.1
## [105] pillar_1.9.0 grid_4.3.0 vctrs_0.6.2 shinystan_2.6.0
## [109] promises_1.2.0.1 arrayhelpers_1.1-0 xtable_1.8-4 evaluate_0.21
## [113] mvtnorm_1.1-3 cli_3.6.1 compiler_4.3.0 rlang_1.1.1
## [117] crayon_1.5.2 rstantools_2.3.1 labeling_0.4.2 ps_1.7.5
## [121] plyr_1.8.8 stringi_1.7.12 rstan_2.21.8 viridisLite_0.4.2
## [125] munsell_0.5.0 colourpicker_1.2.0 Brobdingnag_1.2-9 V8_4.3.0
## [129] fontquiver_0.2.1 Matrix_1.5-4 hms_1.1.3 bit64_4.0.5
## [133] gfonts_0.2.0 shiny_1.7.4 highr_0.10 igraph_1.4.2
## [137] RcppParallel_5.1.7 bslib_0.4.2 bit_4.0.5 officer_0.6.2
## [141] katex_1.4.1 equatags_0.2.0
```

### References

*Estimating survival (time-to-event) models with rstanarm*. https://github.com/stan-dev/rstanarm/blob/feature/frailty-models/vignettes/surv.Rmd

*Bayesian survival analysis using the rstanarm R package*. https://arxiv.org/abs/2002.09633

*Child Development*,

*67*(2), 344–359. https://doi.org/10.2307/1131818

*Journal of the Royal Statistical Society: Series B (Methodological)*,

*34*(2), 187–202. https://doi.org/10.1111/j.2517-6161.1972.tb00899.x

*Applied longitudinal data analysis: Modeling change and event occurrence*. Oxford University Press, USA. https://oxford.universitypressscholarship.com/view/10.1093/acprof:oso/9780195152968.001.0001/acprof-9780195152968

**R**already has a built-in function to convert probabilities to the log-odds scale. Somewhat confusingly, it’s called`qlogis()`

. You can learn more by executing`?qlogis`

or by browsing through this great blog post by Roman Cheplyaka. It’s generally a good idea to stick to the functions in base**R**rather than make your own, like we did earlier in this chapter (see this twitter thread). Since the name of`qlogis()`

isn’t the easiest to remember, you can always execute something like`log_odds <- qlogis`

or`logit <- qlogis`

at the beginning of your scripts and then use one of those as a thin wrapper for`qlogis()`

.↩︎