Binary outcomes

- Most models so far have had response $Y$ as continuous.
- Many responses in practice fall into the $YES/NO$ framework.
- Examples:
- medical: presence or absence of cancer
- financial: bankrupt or solvent
- industrial: passes a quality control test or not

- For $0-1$ responses we need to model $$\pi(x_1, \dots, x_p) = P(Y=1|X_1=x_1,\dots, X_p=x_p)$$
- That is, $Y$ is Bernoulli with a probability that depends on covariates $\{X_1, \dots, X_p\}.$
**Note:**$\text{Var}(Y) = \pi ( 1 - \pi) = E(Y) \cdot ( 1- E(Y))$**Or,**the binary nature forces a relation between mean and variance of $Y$.- This makes logistic regression a
`Generalized Linear Model`

.

- A local health clinic sent fliers to its clients to encourage everyone, but especially older persons at high risk of complications, to get a flu shot in time for protection against an expected flu epidemic.
- In a pilot follow-up study, 50 clients were randomly selected and asked whether they actually received a flu shot. $Y={\tt Shot}$
- In addition, data were collected on their age $X_1={\tt Age}$ and their health awareness $X_2={\tt Health.Aware}$

- Simplest model $\pi(X_1,X_2) = \beta_0 + \beta_1 X_1 + \beta_2 X_2$
- Problems / issues:
- We must have $0 \leq E(Y) = \pi(X_1,X_2) \leq 1$. OLS will not force this.
- Ordinary least squares will not work because of relation between mean and variance.

- Logistic model $\pi(X_1,X_2) = \frac{\exp(\beta_0 + \beta_1 X_1 + \beta_2 X_2)}{1 + \exp(\beta_0 + \beta_1 X_1 + \beta_2 X_2)}$
- This automatically fixes $0 \leq E(Y) = \pi(X_1,X_2) \leq 1$.
**Note:**$\text{logit}(\pi(X_1, X_2)) = \log\left(\frac{\pi(X_1, X_2)}{1 - \pi(X_1,X_2)}\right) = \beta_0 + \beta_1 X_1 + \beta_2 X_2$

In [ ]:

```
g.inv <- function(x) {
return(exp(x) / (1 + exp(x)))
}
x = seq(-4, 4, length=200)
plot(x, g.inv(x), lwd=2, type='l', col='red', cex.lab=2)
```

In [ ]:

```
logit = function(p) {
return(log(p / (1 - p)))
}
p = seq(0.01,0.99,length=200)
plot(p, logit(p), lwd=2, type='l', col='red', cex.lab=2)
```

- Models $E(Y)$ as $F(\beta_0 + \beta_1 X_1 + \beta_2 X_2)$ for some increasing function $F$ (usually a distribution function).
- The logistic model uses the function (we called
`g.inv`

above) $$F(x)=\frac{e^x}{1+e^x}.$$ - Can be fit using Maximum Likelihood / Iteratively Reweighted Least Squares.
For logistic regression, coefficients have nice interpretation in terms of

`odds ratios`

(to be defined shortly).What about inference?

Instead of sum of squares, logistic regression
uses *deviance*:

- $DEV(\mu| Y) = -2 \log L(\mu| Y) + 2 \log L(Y| Y)$ where $\mu$ is a location estimator for $Y$.
- If $Y$ is Gaussian with independent $N(\mu_i,\sigma^2)$ entries then $DEV(\mu| Y) = \frac{1}{\sigma^2}\sum_{i=1}^n(Y_i - \mu_i)^2$
- If $Y$ is a binary vector, with mean vector $\pi$ then $DEV(\pi| Y) = -2 \sum_{i=1}^n \left( Y_i \log(\pi_i) + (1-Y_i) \log(1-\pi_i) \right)$

**Minimizing deviance $\iff$ Maximum Likelihood**

For any binary regression model, $\pi=\pi(\beta)$.

The deviance is: $$\begin{aligned} DEV(\beta| Y) &= -2 \sum_{i=1}^n \left( Y_i {\text{logit}}(\pi_i(\beta)) + \log(1-\pi_i(\beta)) \right) \end{aligned}$$

For the logistic model, the RHS is: $$ \begin{aligned} -2 \left[ (X\beta)^Ty + \sum_{i=1}^n\log \left(1 + \exp \left(\sum_{j=1}^p X_ij \beta_j\right) \right)\right] \end{aligned}$$

The logistic model is special in that $\text{logit}(\pi(\beta))=X\beta$. If we used a different transformation, the first part would not be linear in $X\beta$.

In [ ]:

```
flu.table = read.table('http://stats191.stanford.edu/data/flu.table',
header=TRUE)
flu.glm = glm(Shot ~ Age + Health.Aware, data=flu.table,
family=binomial())
summary(flu.glm)
```

- One reason logistic models are popular is that the parameters have simple interpretations in terms of
**odds**$$ODDS(A) = \frac{P(A)}{1-P(A)}.$$ - Logistic model: $$OR_{X_j} = \frac{ODDS(Y=1|\dots, X_j=x_j+1, \dots)}{ODDS(Y=1|\dots, X_j=x_j, \dots)} = e^{\beta_j}$$
- If $X_j \in {0, 1}$ is dichotomous, then odds for group with $X_j = 1$ are $e^{\beta_j}$ higher, other parameters being equal.

- When incidence is rare, $P(Y=0)\approxeq 1$ no matter what the covariates $X_j$’s are.
- In this case, odds ratios are almost ratios of probabilities: $$OR_{X_j} \approxeq \frac{{\mathbb{P}}(Y=1|\dots, X_j=x_j+1, \dots)}{{\mathbb{P}}(Y=1|\dots, X_j=x_j, \dots)}$$
- Hypothetical example: in a lung cancer study, if $X_j$ is an indicator of smoking or not, a $\beta_j$ of 5 means for smoking vs. non-smoking means smokers are $e^5 \approx 150$ times more likely to develop lung cancer

- In flu example, the odds for a 45 year old with health awareness 50 compared to a 35 year old with the same health awareness are $$e^{-1.429284+3.647052}=9.18$$

In [ ]:

```
logodds = predict(flu.glm, list(Age=c(35,45),Health.Aware=c(50,50)))
logodds
```

The estimated probabilities are below, yielding a ratio of $0.1932/0.0254 \approx 7.61$.

In [ ]:

```
exp(logodds)/(1+exp(logodds))
```

- Initialize $\widehat{\pi}_i = \bar{Y}, 1 \leq i \leq n$
- Define $$Z_i = g(\widehat{\pi}_i) + g'(\widehat{\pi}_i) (Y_i - \widehat{\pi_i})$$
- Fit weighted least squares model $$Z_i \sim \beta_0 + \sum_{j=1}^p \beta_j X_{ij}, \qquad w_i = \widehat{\pi_i} (1 - \widehat{\pi}_i)$$
- Set $\widehat{\pi}_i = \text{logit}^{-1} \left(\widehat{\beta}_0 + \sum_{j=1}^p \widehat{\beta}_j X_{ij}\right)$.
- Repeat steps 2-4 until convergence. This is essentially Newton-Raphson to minimize deviance.

The Newton-Raphson updates for logistic regression are $$ \hat{\beta} \mapsto \hat{\beta} - \nabla^2 DEV(\hat{\beta})^{-1} \nabla DEV(\hat{\beta}) $$

- These turn out to be the same as the updates above!

One thing the IRLS procedure tells us is what the approximate limiting distribution is.

- The IRLS procedure suggests using approximation $\widehat{\beta} \approx N(\beta, (X^TWX)^{-1})$
- This allows us to construct CIs, test linear hypotheses, etc.
- Intervals formed this way are called
*Wald intervals*.

In [ ]:

```
center = coef(flu.glm)['Age']
SE = sqrt(vcov(flu.glm)['Age', 'Age'])
U = center + SE * qnorm(0.975)
L = center + SE * qnorm(0.025)
data.frame(L, center, U)
```

- The estimated covariance uses the weights computed from the fitted model.

In [ ]:

```
pi.hat = fitted(flu.glm)
W.hat = pi.hat * (1 - pi.hat)
X = model.matrix(flu.glm)
C = solve(t(X) %*% (W.hat * X))
c(SE, sqrt(C['Age', 'Age']))
```

The intervals above are slightly different from what `R`

will give you if you ask it for
confidence intervals.

`R`

uses so-called profile intervals.For large samples the two methods should agree quite closely.

In [ ]:

```
confint(flu.glm)
```

What about comparing full and reduced model?

- For a model ${\cal M}$, $DEV({\cal M})$ replaces $SSE({\cal M})$.
- In least squares regression, we use $$\frac{1}{\sigma^2}\left( SSE({\cal M}_R) - SSE({\cal M}_F) \right) \overset{H_0:{\cal M}_R}{\sim} \chi^2_{df_R-df_F}$$
- This is replaced with $$DEV({\cal M}_R) - DEV({\cal M}_F) \overset{n \rightarrow \infty, H_0:{\cal M}_R}{\sim} \chi^2_{df_R-df_F}$$
- Resulting tests do not agree numerically with those coming from IRLS (Wald tests). Both are often used.

In [ ]:

```
anova(glm(Shot ~ Health.Aware, data=flu.table, family=binomial()),
flu.glm)
```

We should compare this difference in deviance with a $\chi^2_1$ random variable.

In [ ]:

```
1 - pchisq(16.863, 1)
```

Let's compare this with the Wald test:

In [ ]:

```
summary(flu.glm)
```

- Similar to least square regression, only residuals used are
*deviance residuals*$r_i = \text{sign}(Y_i-\widehat{\pi}_i) \sqrt{DEV(\widehat{\pi}_i|Y_i)}.$

In [ ]:

```
par(mfrow=c(2,2))
plot(flu.glm)
```

In [ ]:

```
influence.measures(flu.glm)
```

As the model is a likelihood based model, each fitted model has an AIC. Stepwise selection can be used easily …

In [ ]:

```
step(flu.glm, scope=list(upper= ~.^2), direction='both')
```

In [ ]:

```
library(glmnet)
X = model.matrix(flu.glm)[,2:3]
Y = as.numeric(flu.table$Shot)
G = glmnet(X, Y, family="binomial")
plot(G)
```

In [ ]:

```
library(ElemStatLearn)
data(spam)
dim(spam)
X = model.matrix(spam ~ ., data=spam)
Y = as.numeric(spam$spam == 'spam')
G = glmnet(X, Y, family='binomial')
plot(G)
```

In [ ]:

```
CV = cv.glmnet(X, Y, family='binomial')
plot(CV)
c(CV$lambda.min, CV$lambda.1se)
```

- Probit regression model: $$\Phi^{-1}({\mathbb{E}}(Y_i))= \sum_{j=1}^{p} \beta_j X_{ij}$$ where $\Phi$ is CDF of $N(0,1)$, i.e. $\Phi(t) = {\tt pnorm(t)}$.
- Complementary log-log model (cloglog): $$-log(-log({\mathbb{E}}(Y_i)) = \sum_{j=1}^{p} \beta_j X_{ij}.$$
- In logit, probit and cloglog ${\text{Var}}(Y_i)=\pi_i(1-\pi_i)$ but the model for the mean is different.
- Coefficients no longer have an odds ratio interpretation.

In [ ]:

```
summary(glm(Shot ~ Age + Health.Aware, data=flu.table,
family=binomial(link='probit')))
```

Given a dataset $(Y_i, X_{i1}, \dots, X_{ip}), 1 \leq i \leq n$ we consider a model for the distribution of $Y|X_1, \dots, X_p$.

- If $\eta_i=g({\mathbb{E}}(Y_i)) = g(\mu_i) = \beta_0 + \sum_{j=1}^p \beta_j X_{ij}$ then $g$ is called the
*link*function for the model. - If ${\text{Var}}(Y_i) = \phi \cdot V({\mathbb{E}}(Y_i)) = \phi \cdot V(\mu_i)$ for $\phi > 0$ and some function $V$, then $V$ is the called
*variance*function for the model. - Canonical reference Generalized linear models.

- For a logistic model, $g(\mu)={\text{logit}}(\mu), \qquad V(\mu)=\mu(1-\mu).$
- For a probit model, $g(\mu)=\Phi^{-1}(\mu), \qquad V(\mu)=\mu(1-\mu).$
- For a cloglog model, $g(\mu)=-\log(-\log(\mu)), \qquad V(\mu)=\mu(1-\mu).$