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v2.1 notebooks excluding 10,13

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Jonathan Taylor
2023-08-20 19:31:42 -07:00
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Ch11-surv-lab.Rmd
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@@ -1,16 +1,3 @@
---
jupyter:
jupytext:
cell_metadata_filter: -all
formats: ipynb,Rmd
main_language: python
text_representation:
extension: .Rmd
format_name: rmarkdown
format_version: '1.2'
jupytext_version: 1.14.7
---
# Chapter 11
@@ -37,7 +24,7 @@ from ISLP.models import ModelSpec as MS
from ISLP import load_data
```
We also collect the new imports
needed for this lab.
@@ -61,7 +48,7 @@ BrainCancer = load_data('BrainCancer')
BrainCancer.columns
```
The rows index the 88 patients, while the 8 columns contain the predictors and outcome variables.
We first briefly examine the data.
@@ -69,20 +56,20 @@ We first briefly examine the data.
BrainCancer['sex'].value_counts()
```
```{python}
BrainCancer['diagnosis'].value_counts()
```
```{python}
BrainCancer['status'].value_counts()
```
Before beginning an analysis, it is important to know how the
`status` variable has been coded. Most software
uses the convention that a `status` of 1 indicates an
@@ -109,7 +96,7 @@ km_brain = km.fit(BrainCancer['time'], BrainCancer['status'])
km_brain.plot(label='Kaplan Meier estimate', ax=ax)
```
Next we create Kaplan-Meier survival curves that are stratified by
`sex`, in order to reproduce Figure 11.3.
We do this using the `groupby()` method of a dataframe.
@@ -138,7 +125,7 @@ for sex, df in BrainCancer.groupby('sex'):
km_sex.plot(label='Sex=%s' % sex, ax=ax)
```
As discussed in Section 11.4, we can perform a
log-rank test to compare the survival of males to females. We use
the `logrank_test()` function from the `lifelines.statistics` module.
@@ -152,8 +139,8 @@ logrank_test(by_sex['Male']['time'],
by_sex['Female']['status'])
```
The resulting $p$-value is $0.23$, indicating no evidence of a
difference in survival between the two sexes.
@@ -172,7 +159,7 @@ cox_fit = coxph().fit(model_df,
cox_fit.summary[['coef', 'se(coef)', 'p']]
```
The first argument to `fit` should be a data frame containing
at least the event time (the second argument `time` in this case),
as well as an optional censoring variable (the argument `status` in this case).
@@ -186,7 +173,7 @@ with no features as follows:
cox_fit.log_likelihood_ratio_test()
```
Regardless of which test we use, we see that there is no clear
evidence for a difference in survival between males and females. As
we learned in this chapter, the score test from the Cox model is
@@ -206,7 +193,7 @@ fit_all = coxph().fit(all_df,
fit_all.summary[['coef', 'se(coef)', 'p']]
```
The `diagnosis` variable has been coded so that the baseline
corresponds to HG glioma. The results indicate that the risk associated with HG glioma
is more than eight times (i.e. $e^{2.15}=8.62$) the risk associated
@@ -233,7 +220,7 @@ def representative(series):
modal_data = cleaned.apply(representative, axis=0)
```
We make four
copies of the column means and assign the `diagnosis` column to be the four different
diagnoses.
@@ -245,7 +232,7 @@ modal_df['diagnosis'] = levels
modal_df
```
We then construct the model matrix based on the model specification `all_MS` used to fit
the model, and name the rows according to the levels of `diagnosis`.
@@ -272,7 +259,7 @@ fig, ax = subplots(figsize=(8, 8))
predicted_survival.plot(ax=ax);
```
## Publication Data
The `Publication` data presented in Section 11.5.4 can be
@@ -291,7 +278,7 @@ for result, df in Publication.groupby('posres'):
km_result.plot(label='Result=%d' % result, ax=ax)
```
As discussed previously, the $p$-values from fitting Cox’s
proportional hazards model to the `posres` variable are quite
large, providing no evidence of a difference in time-to-publication
@@ -308,8 +295,8 @@ posres_fit = coxph().fit(posres_df,
posres_fit.summary[['coef', 'se(coef)', 'p']]
```
However, the results change dramatically when we include other
predictors in the model. Here we exclude the funding mechanism
variable.
@@ -322,7 +309,7 @@ coxph().fit(model.fit_transform(Publication),
'status').summary[['coef', 'se(coef)', 'p']]
```
We see that there are a number of statistically significant variables,
including whether the trial focused on a clinical endpoint, the impact
of the study, and whether the study had positive or negative results.
@@ -372,7 +359,7 @@ model = MS(['Operators',
intercept=False)
X = model.fit_transform(D)
```
It is worthwhile to take a peek at the model matrix `X`, so
that we can be sure that we understand how the variables have been coded. By default,
the levels of categorical variables are sorted and, as usual, the first column of the one-hot encoding
@@ -382,7 +369,7 @@ of the variable is dropped.
X[:5]
```
Next, we specify the coefficients and the hazard function.
```{python}
@@ -431,7 +418,7 @@ W = np.array([sim_time(l, cum_hazard, rng)
D['Wait time'] = np.clip(W, 0, 1000)
```
We now simulate our censoring variable, for which we assume
90% of calls were answered (`Failed==1`) before the
customer hung up (`Failed==0`).
@@ -443,13 +430,13 @@ D['Failed'] = rng.choice([1, 0],
D[:5]
```
```{python}
D['Failed'].mean()
```
We now plot Kaplan-Meier survival curves. First, we stratify by `Center`.
```{python}
@@ -462,7 +449,7 @@ for center, df in D.groupby('Center'):
ax.set_title("Probability of Still Being on Hold")
```
Next, we stratify by `Time`.
```{python}
@@ -475,7 +462,7 @@ for time, df in D.groupby('Time'):
ax.set_title("Probability of Still Being on Hold")
```
It seems that calls at Call Center B take longer to be answered than
calls at Centers A and C. Similarly, it appears that wait times are
longest in the morning and shortest in the evening hours. We can use a
@@ -488,8 +475,8 @@ multivariate_logrank_test(D['Wait time'],
D['Failed'])
```
Next, we consider the effect of `Time`.
```{python}
@@ -498,8 +485,8 @@ multivariate_logrank_test(D['Wait time'],
D['Failed'])
```
As in the case of a categorical variable with 2 levels, these
results are similar to the likelihood ratio test
from the Cox proportional hazards model. First, we
@@ -514,8 +501,8 @@ F = coxph().fit(X, 'Wait time', 'Failed')
F.log_likelihood_ratio_test()
```
Next, we look at the results for `Time`.
```{python}
@@ -527,8 +514,8 @@ F = coxph().fit(X, 'Wait time', 'Failed')
F.log_likelihood_ratio_test()
```
We find that differences between centers are highly significant, as
are differences between times of day.
@@ -544,8 +531,8 @@ fit_queuing = coxph().fit(
fit_queuing.summary[['coef', 'se(coef)', 'p']]
```
The $p$-values for Center B and evening time
are very small. It is also clear that the
hazard --- that is, the instantaneous risk that a call will be