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OF UNIVERSITY ESSEX INSTITUTE FOR SOCIAL AND ECONOMIC RESEARCH Professor Stephen P. Jenkins <stephenj@essex.ac.uk> Essex Summer School course Survival Analysis and EC968. Part II: Introduction to the analysis of spell duration data
Lesson 7. Unobserved heterogeneity (frailty)
Contents
1 2 AIM ................................................................................................................................................1 INTRODUCTION.........................................................................................................................1
2.1 2.2 2.3
3 4
Model specification........................................................................................1 The implications of unobserved heterogeneity ..............................................4 Frailty models available in Stata overview.................................................4
CONTINUOUS TIME PARAMETRIC MODELS ...................................................................5 DISCRETE TIME MODELS ....................................................................................................13
4.1
5
Prediction for discrete-time frailty models ..................................................18
EXERCISE 7.1 ............................................................................................................................22
1
Aim
The aim of this Lesson is to show how to estimate models incorporating unobserved heterogeneity (or frailty as biostatisticians label it).
2
Introduction
The models considered in this Lesson are a generalisation of those that we considered in Lessons 5 and 6. 2.1 Model specification
For the continuous time parametric models that we estimated in Lesson 5, we now write the hazard rate for each observation as v(t, X) (t, X | v) = (t, X).v where (t, X) is the hazard function considered earlier (and assuming an absence of time-varying covariates for now). Thus unobserved differences between observations are introduced via a multiplicative scaling factor, v. This is a random variable taking 1 Lesson 7
on positive values, with the mean normalised to one (for identification reasons) and finite variance 2. A crucial assumption in these models is that v is distributed independently of X and t. It can be shown that the frailty survivor function is related to the non-frailty one by the relationship: Sv(t, X) S(t, X | v) = [S(t, X)]v. Thus unobserved differences also imply a scaling of the non-frailty survivor function. Observe that, for proportional hazards models, the frailty hazard rate may be written as v(t, X) (t, X | , v) = 0(t).exp(X).v = 0(t).exp(X + u) or ln[v(t, X)] = ln[0(t)] + X + u where 0(t) is the baseline hazard function and the error term u ln(v) which is random variable with a mean of zero. The random variable v, or equivalently u, may be interpreted in several ways. The most common one is that it summarises the impact of omitted variables on the hazard rate whether the missing regressors are intrinsically unobservable or simply unobserved in the data set to hand. Alternative interpretations can be offered in terms of errors of measurement in recorded regressors or recorded survival times (see Lancaster 1990, Chapter 4). In the discrete time proportional hazards model, the model specification follows directly from above. The standard cloglog model generalises to: cloglog[p(t, X | , v)] = D(t) + X + u where D(t) characterises the baseline hazard function. The logistic hazard regression model is typically generalised in an analogous way: logit[p(t, X | , v)] = D(t) + X + e where the error term e is a random variable with mean zero and finite variance. These are random intercept models where randomness is characterized using some parametric distribution (see below). To estimate these models requires expressions for survival and density functions that do not condition on the unobserved effects for, since each individual v is unobserved, how could one write down the likelihood contribution for each observation? The way forward to specify a distribution for the v, where the distribution is characterised in terms of parameters (that can be estimated), and the unconditional survivor function is written in terms of this. This is known as integrating out the unobserved effect. Referring to the example above, one works with survivor function S(t, X | , 2) rather than S(t, X | , v), and similarly for the density function. In principle, any continuous distribution with positive support, mean one and finite variance, is a suitable candidate to represent the frailty distribution. For tractability reasons, however, the choice of distribution is typically limited to those that provide a closed form expression for the frailty survivor function.
2
Lesson 7
For continuous time models the Gamma and Inverse Gaussian distributions have been the two that have been most commonly used. For the Gamma mixture model, the survivor function is given by S(t, X | , V) = ( 1 V.ln[S(t)] )1/V where V 2 and the non-frailty survivor function is S(t). For the Weibull model, ln[S(t)] = t where = exp(X), and so in this case, S(t, X | , V) = ( 1 + V.t )1/V and the median duration (integrating over the distribution of the frailty v) is [(2V 1)/(V)]1/. One could derive predicted survival probabilities assuming that v took on a particular value, for example the mean v = 1. In this case, the formula for the median is the same as the standard non-frailty median formula, but of course evaluated using the parameters estimated from the model with unobserved heterogeneity. For the Inverse Gaussian mixture model, the survivor function is given by S(t, X | , V) = exp[ (1/V)(1 {1 2V.ln[S(t)]}1/2) ]. For the Weibull model, the median duration (averaging over the frailty v) is [{ [1 + V.ln(2)]2 1}/(2V)]1/. Estimates of the median conditioning on particular frailty v values can also be derived, as for the Gamma model. For the discrete time PH model, the Gamma distribution has been the most popular distribution. For cloglog and logistic models, it also straightforward to assume a Normal (Gaussian) distribution for u and e, respectively. (In these latter two cases, closed form expressions are not available; numerical quadrature techniques are used for the integrating out.) Prediction of survivor functions is rather more complicated than for the continuous time case (as Lesson 6 showed), as the survivor function is a product of the complements of the period-specific hazard rates. And, with frailty, these hazard rates also depend on an unobserved individual error term. The most common empirical practice has been to calculate survivor functions conditioning on a particular error term value using the estimates of covariate coefficients from the frailty model but setting the error term equal to its mean. There is also a literature, following the pioneering work of Heckman and Singer in the 1980s, which eschewed parametric forms for the frailty distribution. Instead nonparametric characterisations were proposed, by which an arbitrary distribution was fitted using a set of parameters representing a set of mass points along the distributions support together with the probabilities of a subject being at each mass point. Consider for illustrative purposes a discrete time proportional hazards model. When there is no frailty, the discrete hazard rate in period t is ht = 1 exp(exp(0 + Xit)) where 0 is an intercept and the linear index function, Xit, incorporates the impact of covariates Xit. Suppose now that each individual belongs to one of a number of different types, and membership of each class is unobserved. This is parameterized by allowing the intercept term in the hazard function to differ across types. For example, for a model with types z = 1, ..., Z, the hazard function for an individual belonging to type z is: hzt = 1 exp(exp(mz + 0 + Xit)) and the probability of belonging to type z is pz. The mz characterize the discrete points of support of a multinomial distribution (mass points), with m1 normalized to equal zero and p1 = 1 z = 2,,Z (pz). Mass point z equals mz + 0. This is a random intercept model where randomness is characterized using a discrete distribution.
3
Lesson 7
We focus mostly on parametric representations of unobserved heterogeneity in this lesson. The particular models considered are listed in the overview provided by Section 2.3 below. 2.2 The implications of unobserved heterogeneity
What happens to parameter estimates if one (mistakenly) ignores unobserved heterogeneity? The theoretical literature has suggested several results, typically derived with reference to a continuous time PH model: The non-frailty model will over-estimate the degree of negative duration dependence in the (true) baseline hazard, and under-estimate the degree of positive duration dependence. (This is a selection effect. In the negative duration dependence case, observations with high v values fail faster, other things equal, so the survivors at any given survival time are increasingly composed of observations with relatively low v values and thence lower hazard rates.) The proportionate effect of a given regressor on the hazard rate is no longer constant and independent of survival time (in the non-frailty PH model, the proportionate effect for regressor Xk is the fixed amount k). The presence of unobserved heterogeneity attentuates the proportionate response of the hazard to variation in each regressor at any survival time. In short the estimate of a positive (negative) k derived from the (wrong) no-frailty model will underestimate (overrestimate) the true estimate. (Lancaster, 1990, chapter 4, proves this for the case when v follows a Gamma distribution.) The empirical literature has generally confirmed these results. There has also been discussion of the magnitude of the effects and how serious the biases are in practice. Verdicts have been contingent on the choice of shape of the non-frailty hazard function and the choice of the distribution for the unobserved heterogeneity. The results from several recent papers have suggested that if a fully flexible specification for the baseline hazard function is used, then the magnitude of the biases in the nonfrailty model (relative to the true model) are diminished. In sum, the literature to date provides a number of important results and guidelines, but conclusions about the empirical relevance of unobserved heterogeneity are likely to differ from application to application. Moreover, frailty models can be relatively fragile in the statistical sense they can be relatively hard to fit especially if the frailty variance is close to zero. 2.3 Frailty models available in Stata overview
For continuous time models, Stata estimates frailty generalisations of all the nonfrailty parametric models that were cited in Lesson 5: Exponential, Weibull, Gompertz, Log-logistic, Lognormal, Gamma. As we shall see below, estimation is in principle as straightforward as adding a frailty(.) option to ones streg command, and choosing between Gamma and Inverse Gaussian representations of the frailty. I say in principle because the frailty models can sometimes be difficult to fit.
4
Lesson 7
Several discrete time survival models with frailty can be estimated in Stata. The discrete time PH (cloglog) model with Gamma heterogeneity can be estimated using my program pgmhaz8. This can be downloaded for free from the SSC-IDEAS archive using the command ssc install pgmhaz8. The cloglog model with Normal distributed errors can be estimated using Statas xtcloglog command and the logistic model with Normal distributed errors can be estimated using xtlogit. (Note that a Normal distribution for u corresponds to a lognormal distribution for v.) Illustrations of the programs cited are provided below. Section 3 considers continuous time models and Section 4 discrete time models. Some discrete time models with Heckman and Singer-type non-parametric representations of frailty can be estimated using my program hshaz (for proportional hazard models, obtained via ssc install hshaz), or Sophia Rabe-Heskeths program gllamm (obtained via ssc install gllamm). Split-population (or cure) survival models can also be interpreted as models with a particular type of mover-stayer unobserved heterogeneity. See e.g. my program for discrete time data, spsurv. This is downloadable using ssc install spsurv. Two types of frailty are currently incorporated in the Stata streg programs. The first (and default) is known as observation level frailty there is one value of v for each record in the data. This is what we focus on here. If there is multiple record data, e.g. because of episode splitting to incorporate time-varying covariates, then each separate record counts as an observation. This treatment of frailty may not be what you want. For example you may wish to have a single value of v that is common to a group of observations (e.g. the same individual or the same family). This is known as shared frailty. To incorporate this you need to specify the shared option in addition to the frailty(.) one. If there is only a single record per person (such as when there are no time-varying covariates), or there is non-informative episode splitting, then the two approaches are equivalent. The stcox command for Cox PL regression includes an option for estimating models with shared frailty, assuming a Gamma mixture. In the illustrative data that we use here, there is only a single record per person (see previous paragraph), and so the command is not applicable. The discrete time frailty models cited above all incorporate shared frailty rather than observation level frailty.
3
Continuous time parametric models
Let us now illustrate the frailty models available via streg. The discussion here draws heavily on that in the Stata 7 Reference Manual Volume 3, QSt, pp. 359363, and uses the same data set, the breast cancer data (bc.dta). These (hypothetical) data refer to survival times for 80 women with breast cancer. Covariates summarise patients age, whether she smokes, and average weekly calorific intake over the course of the study.
5
Lesson 7
We shall first estimate a Weibull model without frailty but using all the covariates. (The data were in fact created by simulation from a Weibull distribution.)
. use bc.dta, clear . . stset t, f(dead) failure event: obs. time interval: exit on or before: dead ~= 0 & dead ~= . (0, t] failure
-----------------------------------------------------------------------------80 total obs. 0 exclusions -----------------------------------------------------------------------------80 obs. remaining, representing 58 failures in single record/single failure data 1257.07 total analysis time at risk, at risk from t = 0 earliest observed entry t = 0 last observed exit t = 35 . streg age smoking dietfat, d(weib) nohr nolog failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -13.352142 Number of obs = 80
LR chi2(3) Prob > chi2
= =
250.96 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .559197 .0563239 9.93 0.000 .4488042 .6695899 smoking | 1.649311 .3276501 5.03 0.000 1.007128 2.291493 dietfat | 2.222411 .2404553 9.24 0.000 1.751128 2.693695 _cons | -45.97988 4.634153 -9.92 0.000 -55.06265 -36.89711 -------------+---------------------------------------------------------------/ln_p | 1.431728 .0978872 14.63 0.000 1.239872 1.623583 -------------+---------------------------------------------------------------p | 4.185925 .4097485 3.455172 5.071228 1/p | .2388958 .0233848 .1971909 .2894212 ------------------------------------------------------------------------------
We can see that higher hazard rates shorter survival times are positively associated with age, smoking, and dietary fat at conventional levels of statistical significance. Let us now drop the dietfat variable with the aim of introducing unobserved heterogeneity. We will use this next model as the reference non-frailty model:
6
Lesson 7
. streg age smoking , d(weib) nohr nolog failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -79.419727 Number of obs = 80
LR chi2(2) Prob > chi2
= =
118.82 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .1644213 .0149837 10.97 0.000 .1350538 .1937888 smoking | .9056537 .3061656 2.96 0.003 .3055801 1.505727 _cons | -11.20242 .9989083 -11.21 0.000 -13.16024 -9.244594 -------------+---------------------------------------------------------------/ln_p | .3633523 .0955797 3.80 0.000 .1760195 .5506852 -------------+---------------------------------------------------------------p | 1.438142 .1374573 1.192461 1.734441 1/p | .6953414 .0664605 .5765546 .8386016 ------------------------------------------------------------------------------
Let us also predict median survival times. First we predict the median for the person with characteristics equal to the sample mean values on all covariates, and then we predict the median for each person in the sample. The first prediction has to be done manually but the second can be done directly using predict.
. predict xb, xb . su xb Variable | Obs Mean Std. Dev. Min Max -------------+----------------------------------------------------xb | 80 -3.834933 1.93139 -6.927465 -.1026448 . di "Pred. Median [at sample mean X] = " (ln(2)*exp(-r(mean)))^(1/e(aux_p)) Pred. Median [at sample mean X] = 11.153305 . * median duration for each person in sample . * NB Stata allows you to generate these directly: . predict mediand, time (option median time assumed; predicted median time) . su mediand, de predicted median _t ------------------------------------------------------------Percentiles Smallest 1% .8323699 .8323699 5% 1.109575 1.046216 10% 1.707746 1.046216 Obs 80 25% 3.789622 1.046216 Sum of Wgt. 80 50% 75% 90% 95% 99% 11.5727 34.23132 67.97328 80.82134 95.78455 Largest 85.43642 95.78455 95.78455 95.78455 Mean Std. Dev. Variance Skewness Kurtosis 23.37674 26.22382 687.6885 1.359739 3.853934
. drop xb mediand
So the median duration for the person with mean characteristics is 11.2, and the median among the sample as a whole is 11.6.
7
Lesson 7
Now we allow for unobserved heterogeneity, first assuming a Gamma mixture distribution and then an Inverse Gaussian one. We shall also look at predicted median distributions.
. streg age smoking , d(weib) nohr nolog frailty(gamma) failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form Gamma frailty No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -68.135804 Number of obs = 80
LR chi2(2) Prob > chi2
= =
135.75 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .3893002 .0934984 4.16 0.000 .2060467 .5725537 smoking | 1.025521 .5225054 1.96 0.050 .0014291 2.049613 _cons | -23.8082 5.204923 -4.57 0.000 -34.00966 -13.60674 -------------+---------------------------------------------------------------/ln_p | 1.087761 .222261 4.89 0.000 .6521376 1.523385 /ln_the | .3307466 .5250758 0.63 0.529 -.698383 1.359876 -------------+---------------------------------------------------------------p | 2.967622 .6595867 1.91964 4.587727 1/p | .3369701 .0748953 .2179729 .520931 theta | 1.392007 .7309092 .4973889 3.895711 -----------------------------------------------------------------------------Likelihood ratio test of theta=0: chibar2(01) = 22.57 Prob>=chibar2 = 0.000
The theta value reported in the output is the estimate of the frailty distribution variance. Note that the frailty model is preferred to the reference non-frailty model according to the relevant likelihood ratio test. The test is a boundary test that takes account of the fact that the null distribution is not the usual chi-squared(d.f. = 1) but is rather a 50:50 mixture of a chi-squared(d.f. = 0) variate (which is a point mass at zero) and chi-squared(d.f. = 1) hence the reference to chibar2(01) in the output. Click on the blue chibar2(01) in the output window for an explanation, or see Gutierrez et al. (2001) for more details (Gutierrez, R.G., Carter, S., and Drukker, D., On boundaryvalue likelihood-ratio tests, insert sg160, Stata Technical Bulletin, STB-60, StataCorp, College Station TX.) The p-value = 0.000 in this case. The frailty has expected effects on model parameters. The estimated coefficients on the regressors age and smoking are larger in magnitude that the corresponding coefficients in the reference model. Also the Weibull distribution shape parameter p is larger in the frailty models than in the reference model the baseline hazard slopes upwards to a greater extent. Observe too that with frailty present, an exponentiated coefficient is simply that, losing its interpretation in terms of a hazard ratio (a proportional change in the hazard for a one unit change in the relevant covariate). Now let us consider the predictions of median duration and compare them with those of the reference non-frailty model. Observe that because there are no time-varying covariates, observation-level and shared frailty models are equivalent. The former is the default.
8
Lesson 7
. predict xb, xb (option unconditional assumed) . su xb Variable | Obs Mean Std. Dev. Min Max -------------+----------------------------------------------------xb | 80 -6.685866 4.461222 -13.6864 1.353928 . di "Pred. Median [at sample mean (e(theta)*exp(r(mean))))^(1/e(aux_p)) Pred. Median [at sample mean X] = 10.023875 X] = " ((2^e(theta) 1)/
. * median duration for each person in sample . * NB Stata allows you to generate these directly: . predict mediand, time (option unconditional assumed) (option median time assumed; predicted median time) . su mediand, de predicted _t ------------------------------------------------------------Percentiles Smallest 1% .6675103 .6675103 5% .867764 .8271137 10% 1.107688 .8271137 Obs 80 25% 3.071027 .867764 Sum of Wgt. 80 50% 75% 90% 95% 99% 10.94738 37.13202 71.5497 87.29696 106.0531 Largest 93.01468 106.0531 106.0531 106.0531 Mean Std. Dev. Variance Skewness Kurtosis 24.43518 29.05888 844.4185 1.378743 3.908608
The median for the person with mean characteristics is now predicted to be 10.02. This is smaller than the corresponding value for the non-frailty model as expected, perhaps, given the change in the coefficients and increase in duration dependence parameter. We can also see what the median is for the case in which we condition on frailty v = 1 (the mean value):
. predict mediand2, time alpha1 (option median time assumed; predicted median time) . su mediand2, de predicted median _t ------------------------------------------------------------Percentiles Smallest 1% .5600455 .5600455 5% .7280598 .6939539 10% .9293576 .6939539 Obs 80 25% 2.576612 .7280598 Sum of Wgt. 80 50% 75% 90% 95% 99% 9.184921 31.15401 60.03067 73.24273 88.97925 Largest 78.03993 88.97925 88.97925 88.97925 Mean Std. Dev. Variance Skewness Kurtosis 20.50128 24.38059 594.4133 1.378743 3.908608
. . drop mediand mediand2 xb
It turns out that the median is a little bit smaller than before, at all corresponding points of the sample distribution.
9
Lesson 7
Now let us repeat the analysis for the Inverse Gaussian frailty model.
. streg age smoking , d(weib) nohr nolog frailty(invgauss) failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form Inverse-Gaussian frailty No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -73.838578 Number of obs = 80
LR chi2(2) Prob > chi2
= =
125.44 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .2500841 .0360754 6.93 0.000 .1793777 .3207906 smoking | 1.066574 .4311907 2.47 0.013 .2214561 1.911692 _cons | -16.01035 2.097075 -7.63 0.000 -20.12054 -11.90015 -------------+---------------------------------------------------------------/ln_p | .7173904 .1434382 5.00 0.000 .4362567 .9985241 /ln_the | .2374778 .8568064 0.28 0.782 -1.441832 1.916788 -------------+---------------------------------------------------------------p | 2.049079 .2939162 1.546906 2.714273 1/p | .4880241 .0700013 .3684228 .6464518 theta | 1.268047 1.086471 .2364941 6.799082 -----------------------------------------------------------------------------Likelihood ratio test of theta=0: chibar2(01) = 11.16 Prob>=chibar2 = 0.000
It turns out that frailty is again statistically significant, and again the parameters of the model change in the expected direction. The improvement in log-likelihood relative to the no-frailty model is largest for the Gamma mixture model, but choice between the Gamma and Inverse Gaussian specifications is complicated by the fact that the two models are non-nested. How do model predictions differ between the two frailty specifications? We explore this using predict after the estimation command.
10
Lesson 7
. predict xb, xb (option unconditional assumed) . su xb Variable | Obs Mean Std. Dev. Min Max -------------+----------------------------------------------------xb | 80 -4.893819 2.902277 -9.508158 .5614455 . di "Pred. Median [at sample mean 1)/(2*e(theta)*exp(r(mean))) )^(1/e( > aux_p)) Pred. Median [at sample mean X] = 10.883087 = X] " ( ((1+e(theta)*ln(2))^2 -
. * median duration for each person in sample . * NB Stata allows you to generate these directly: . predict mediand, time (option unconditional assumed) (option median time assumed; predicted median time) . su mediand, de predicted _t ------------------------------------------------------------Percentiles Smallest 1% .7595033 .7595033 5% 1.032402 .9694791 10% 1.454846 .9694791 Obs 80 25% 3.711818 .9694791 Sum of Wgt. 80 50% 75% 90% 95% 99% 11.31624 36.10971 71.73492 86.30687 103.4532 Largest 91.56709 103.4532 103.4532 103.4532 Mean Std. Dev. Variance Skewness Kurtosis 24.43678 28.26338 798.8188 1.382187 3.930592
. drop xb mediand
The predicted median for the person with average characteristics is now 10.9. What happens if we re-estimate the models but now re-introduce dietfat as a regressor? Here are the estimates from the model with Gamma frailty.
11
Lesson 7
. streg age smoking dietfat, d(weib) nolog frailty(gamma) nohr failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form Gamma frailty No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -13.352142 Number of obs = 80
LR chi2(3) Prob > chi2
= =
245.32 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .5592066 .0563201 9.93 0.000 .4488212 .6695919 smoking | 1.649354 .3276412 5.03 0.000 1.007188 2.291519 dietfat | 2.222451 .2404404 9.24 0.000 1.751196 2.693706 _cons | -45.98067 4.633832 -9.92 0.000 -55.06281 -36.89852 -------------+---------------------------------------------------------------/ln_p | 1.431747 .0978783 14.63 0.000 1.239909 1.623585 /ln_the | -15.92927 6628.419 -0.00 0.998 -13007.39 12975.53 -------------+---------------------------------------------------------------p | 4.186005 .4097189 3.455299 5.071237 1/p | .2388912 .0233823 .1971906 .2894106 theta | 1.21e-07 .0008006 0 . -----------------------------------------------------------------------------Likelihood ratio test of theta=0: chibar2(01) = 0.00 Prob>=chibar2 = 1.000
Here are the estimates from the model with Inverse Gaussian frailty.
. streg age smoking dietfat, d(weib) nolog frailty(invg) nohr failure _d: analysis time _t: dead t
Weibull regression -- log relative-hazard form Inverse-Gaussian frailty No. of subjects = No. of failures = Time at risk = Log likelihood = 80 58 1257.07 -13.352142 Number of obs = 80
LR chi2(3) Prob > chi2
= =
246.41 0.0000
-----------------------------------------------------------------------------_t | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------age | .5592044 .0563229 9.93 0.000 .4488136 .6695952 smoking | 1.649341 .3276494 5.03 0.000 1.00716 2.291522 dietfat | 2.222442 .2404515 9.24 0.000 1.751166 2.693718 _cons | -45.98049 4.634064 -9.92 0.000 -55.06309 -36.89789 -------------+---------------------------------------------------------------/ln_p | 1.431742 .0978845 14.63 0.000 1.239892 1.623592 /ln_the | -14.424 2866.444 -0.01 0.996 -5632.551 5603.703 -------------+---------------------------------------------------------------p | 4.185987 .4097431 3.455242 5.071276 1/p | .2388923 .0233838 .197189 .2894154 theta | 5.44e-07 .0015598 0 . -----------------------------------------------------------------------------Likelihood ratio test of theta=0: chibar2(01) = 0.00 Prob>=chibar2 = 1.000
For both models, we now see that there is negligible unobserved heterogeneity observe the near-zero frailty variances, and the p-values for the likelihood ratio test equal to one. The coefficients on the covariates are almost exactly the same as those in the corresponding model without unobserved heterogeneity that we estimated at the very start.
12
Lesson 7
4
Discrete time models
We will now repeat our analysis of the same data but use discrete time models instead. To facilitate comparability with the Weibull models of Section 3, we shall use proportional hazards models with a log(time) specification for duration dependence. One initial minor complication is that survival times in bc.dta are non-integer. We therefore need to create a new survival time variable (td below, rather than t), in addition to doing the usual episode-splitting to form a data set organised in personmonth form. We use the ceil() function to round up the survival times to the nearest integer.
use bc, clear . * convert survival times to discrete integers . ge td = ceil(t) // has same effect as: ge td = round(t+.49,1) . su t td Variable | Obs Mean Std. Dev. Min Max -------------+----------------------------------------------------t | 80 15.71337 13.59278 .33 35 td | 80 16.0875 13.40243 1 35 . . sort t . . ge id = _n
Now we do the episode-splitting to derive the data organised in person-month form, and create the appropriate spell month identifier. Then we create the variable to summarize duration dependence in the discrete hazard, log(time) in this case.
. expand td /* expand on td (not t) since time intervals indexed on td */ (1207 observations created) . sort id . by id: ge newt = _n . by id: ge died = dead==1 & _n==_N . ge logt = ln(newt)
Let us begin with the model including all covariates (but no frailty) and then the reference model from which the dietfat regressor has been omitted:
13
Lesson 7
. cloglog died logt age smoking dietfat, nolog Complementary log-log regression Number of obs Zero outcomes Nonzero outcomes LR chi2(4) Prob > chi2 = = = = = 1287 1229 58 244.53 0.0000
Log likelihood = -114.18864
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------logt | 3.594979 .4918103 7.31 0.000 2.631048 4.558909 age | .5449406 .060167 9.06 0.000 .4270153 .6628658 smoking | 1.638842 .3753424 4.37 0.000 .9031841 2.374499 dietfat | 2.149625 .2561884 8.39 0.000 1.647505 2.651745 _cons | -44.72472 4.9638 -9.01 0.000 -54.45359 -34.99585 -----------------------------------------------------------------------------. cloglog died logt age smoking, nolog Complementary log-log regression Number of obs Zero outcomes Nonzero outcomes LR chi2(3) Prob > chi2 = = = = = 1287 1229 58 126.56 0.0000
Log likelihood = -173.17144
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------logt | .5328684 .1743352 3.06 0.002 .1911776 .8745592 age | .1647403 .015838 10.40 0.000 .1336985 .1957821 smoking | .8752191 .3078178 2.84 0.004 .2719074 1.478531 _cons | -11.10694 1.042009 -10.66 0.000 -13.14924 -9.064643 ------------------------------------------------------------------------------
The estimates are similar to the corresponding Weibull model estimates (as expected). Now what if we suppose that the frailty term u is Normally distributed? We use the xtcloglog command. xt stands for cross-section time series or panel data estimator we are using a panel data estimator to estimate a survival analysis model. The mandatory i(.) option is used to identify the observations with distinct values for the heterogeneity term. The sigma_u reported is the standard deviation of the heterogeneity variance. The reported rho is the ratio of the heterogeneity variance to one plus the heterogeneity variance. So if the hypothesis that rho is zero cannot be rejected, then frailty is unimportant.
14
Lesson 7
xtcloglog died logt age smoking , nolog i(id) Random-effects complementary log-log Group variable (i) : id Random effects u_i ~ Gaussian Number of obs Number of groups = = 1287 80 1 16.1 35 21.88 0.0001
Obs per group: min = avg = max = Wald chi2(3) Prob > chi2 = =
Log likelihood
= -164.02912
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------logt | 3.344268 .9695804 3.45 0.001 1.443925 5.244611 age | .5170562 .1222468 4.23 0.000 .2774569 .7566554 smoking | 1.587545 .6984732 2.27 0.023 .2185631 2.956528 _cons | -32.61654 7.741936 -4.21 0.000 -47.79045 -17.44262 -------------+---------------------------------------------------------------/lnsig2u | 1.743016 .5665486 .6326016 2.853431 -------------+---------------------------------------------------------------sigma_u | 2.390514 .677171 1.372043 4.164997 rho | .8510698 .07181 .6530791 .9454958 -----------------------------------------------------------------------------Likelihood ratio test of rho=0: chibar2(01) = 18.28 Prob >= chibar2 = 0.000
The likelihood ratio test suggests statistically significant frailty. The frailty has expected effects on model parameters. The estimated coefficients on the regressors age and smoking are larger in magnitude that the corresponding coefficients in the reference model. Also the duration dependence is much larger in the frailty models than in the reference model the baseline hazard slopes upwards to a greater extent. What happens if we re-estimate our model but instead assume Gamma-distributed unobserved heterogeneity? To investigate this, we use pgmhaz8. Its syntax differs from that for the other programs (it was modelled on the syntax for the Cox model in Stata 5, i.e. in a pre-stset era), but is relatively straightforward. One specifies the covariates after the program name, and then options are used to specify identifiers for each person (the id(.) option), the spell interval identifier for each person (seq(.), which is an integer sequence from 1 to the total survival time for that person) and the censoring variable (the dead(.) option). See help pgmhaz8 for further details. The program first reports the non-frailty cloglog model (estimated using glm), and then the Gamma frailty model. (Observe that the non-frailty estimates correspond to those reported earlier. pgmhaz8 has an option to turn off the reporting of these.)
15
Lesson 7
. pgmhaz8 logt age smoking , id(id) seq(newt) dead(died) nolog PGM hazard model without gamma frailty Generalized linear models Optimization : ML: Newton-Raphson Deviance Pearson = = 346.3428729 827.8333299 No. of obs Residual df Scale parameter (1/df) Deviance (1/df) Pearson = = = = = 1287 1283 1 .2699477 .6452325
Variance function: V(u) = u*(1-u) Link function : g(u) = ln(-ln(1-u)) Standard errors : OIM Log likelihood BIC = -173.1714365 = -8840.02592
[Bernoulli] [Complementary log-log]
AIC
=
.2753247
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------logt | .5328684 .1743352 3.06 0.002 .1911776 .8745592 age | .1647403 .015838 10.40 0.000 .1336985 .1957821 smoking | .8752191 .3078178 2.84 0.004 .2719074 1.478531 _cons | -11.10694 1.042009 -10.66 0.000 -13.14924 -9.064643 -----------------------------------------------------------------------------PGM hazard model with gamma frailty Log likelihood = -162.55447 Number of obs LR chi2() Prob > chi2 = = = 1287 . .
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------hazard | logt | 2.742215 .9879201 2.78 0.006 .8059276 4.678503 age | .4369005 .1197364 3.65 0.000 .2022216 .6715795 smoking | .9551981 .5663482 1.69 0.092 -.154824 2.06522 _cons | -26.43864 6.772158 -3.90 0.000 -39.71182 -13.16545 -------------+---------------------------------------------------------------ln_varg | _cons | .5409866 .5287288 1.02 0.306 -.4953028 1.577276 -------------+---------------------------------------------------------------Gamma var. | 1.717701 .9081978 1.89 0.059 .6093864 4.841749 -----------------------------------------------------------------------------LR test of Gamma var. = 0: chibar2(01) = 21.2339 Prob.>=chibar2 = 2.0e-06
The p-value for the likelihood ratio test is virtually zero, indicating statistically significant frailty which is entirely consistent with what we found for the corresponding continuous time model. We see again too that regressor coefficients are larger in magnitude and the degree of positive duration dependence in the hazard is larger. If we add dietfat back into the model as a regressor, then the frailty appears to become negligible. Observe that I omitted the nolog option this time, so the iteration log for the frailty model is reported. There are obviously some problems with convergence (note the not concave messages) and, although the model did finally converge satisfactorily, the frailty variance is tiny.
16
Lesson 7
. pgmhaz8 logt age smoking dietfat, id(id) seq(newt) dead(died) nolog . pgmhaz8 logt age smoking dietfat, id(id) seq(newt) dead(died) iter(25) PGM hazard model without gamma frailty Generalized linear models Optimization : ML: Newton-Raphson Deviance Pearson = = 228.37728 679.2477092 No. of obs Residual df Scale parameter (1/df) Deviance (1/df) Pearson = = = = = 1287 1282 1 .1781414 .5298344
Variance function: V(u) = u*(1-u) Link function : g(u) = ln(-ln(1-u)) Standard errors : OIM Log likelihood BIC = -114.18864 = -8950.831444
[Bernoulli] [Complementary log-log]
AIC
=
.1852193
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------logt | 3.594979 .4918101 7.31 0.000 2.631049 4.558909 age | .5449406 .060167 9.06 0.000 .4270154 .6628657 smoking | 1.638842 .3753423 4.37 0.000 .9031842 2.374499 dietfat | 2.149625 .2561882 8.39 0.000 1.647506 2.651745 _cons | -44.72472 4.963796 -9.01 0.000 -54.45358 -34.99586 -----------------------------------------------------------------------------Iteration 0: log likelihood = -116.47328 Iteration 1: log likelihood = -114.39827 Iteration 2: log likelihood = -114.29907 Iteration 3: log likelihood = -114.22669 Iteration 4: log likelihood = -114.20848 Iteration 5: log likelihood = -114.19356 Iteration 6: log likelihood = -114.1906 Iteration 7: log likelihood = -114.1895 Iteration 8: log likelihood = -114.18888 Iteration 9: log likelihood = -114.18871 Iteration 10: log likelihood = -114.18865 Iteration 11: log likelihood = -114.18864 numerical derivatives are approximate nearby values are missing Iteration 12: log likelihood = -114.18864 Iteration 13: log likelihood = -114.18864 Iteration 14: log likelihood = -114.18864 PGM hazard model with gamma frailty Log likelihood = -114.18864
(not concave)
(not concave) (not concave)
Number of obs LR chi2() Prob > chi2
= = =
1287 . .
-----------------------------------------------------------------------------died | Coef. Std. Err. z P>|z| [95% Conf. Interval] -------------+---------------------------------------------------------------hazard | logt | 3.595206 .1925172 18.67 0.000 3.217879 3.972532 age | .5449706 .0197679 27.57 0.000 .5062262 .5837149 smoking | 1.638924 .3370305 4.86 0.000 .9783566 2.299492 dietfat | 2.149746 .1156383 18.59 0.000 1.9231 2.376393 _cons | -44.72717 1.595761 -28.03 0.000 -47.85481 -41.59954 -------------+---------------------------------------------------------------ln_varg | _cons | -13.22453 1390.536 -0.01 0.992 -2738.625 2712.176 -------------+---------------------------------------------------------------Gamma var. | 1.81e-06 .002511 0.00 0.999 0 . -----------------------------------------------------------------------------LR test of Gamma var. = 0: chibar2(01) = -8.2e-07 Prob.>=chibar2 = .5
What is probably happening is that the program is trying to find ever-smaller values of the variance: a value of the log of the gamma variance of 13.22 implies a value for the variance that is very very close to zero! But the model is programmed with the gamma variance constrained to be positive hence convergence problems. This type of result is quite common when estimating frailty models.
17
Lesson 7
To reiterate, although our first applications of pgmhaz8 converged relatively quickly and without difficulty, this is often not the case in practice with real-world data sets. With large data sets, the model can be slow to converge, because the program uses numerical derivatives. Moreover the likelihood surface is not globally concave and non-concavities may sometimes be reported at the final iteration, or the maximisation may sometimes get stuck on a flat part and fail to converge with an error message:
pgmhaz8_ll does not compute a continuous nonconstant function could not calculate numerical derivatives r(430);
If such situations arise, users are recommended to use the trace option to pgmhaz8 e.g. check whether the estimate of ln(Gamma variance) is heading off towards minus infinity and to experiment with different starting values for the Gamma variance using the lnvar(.) option. xtcloglog and xtlogit may also take a long time to run with expanded person-month data sets if they are large. 4.1 Prediction for discrete-time frailty models
There are no built-in commands for predicting medians and so on for the discrete-time frailty models, as there were for the frailty models estimated using streg. However predictions can be derived straightforwardly, by adapting the strategy that was illustrated in Lesson 6 for discrete-time models without unobserved heterogeneity. For the discrete-time frailty models, predictions are derived assuming that the frailty term is set equal to its mean value. The idea is to first derive predicted hazard rates for persons with given characteristics, and thence the implied survivor functions. I start with the pgmhaz8 model estimated earlier, with age and smoking as the covariates, and consider predictions for three persons: (a) someone with the sample mean values of each regressor (age = 35, smoking = 0.23); (b) someone aged 30 and a non-smoker; and (c) someone aged 50 and a smoker. Recall that the cloglog hazard rate with frailty has specification: p(t) = 1 exp[exp( z(t) )] where z(t) = D(t) + X + u. D(t) is the baseline hazard function and X includes an intercept term. For predictions, we will take the case with u = 0, and we will consider the case with D(t) = (q1)log(t), as earlier. What we have to do is, first, generate values of z(t) for each of our three persons, using the coefficient estimates left behind after pgmhaz8 Model 2, and then, second, the implied hazard rate and survivor functions. At the first step, one uses Statas matrix operations to extract the required coefficients (one doesnt need the estimate of the frailty variance). Then one generates the values of z(t) = 0 + 1*age + 2*smoking + c.log(t) for the given values of age and smoking. Observe the result used to get z(t) when age and smoking are at sample mean values: (the mean of a linear combination equals the linear combination of means). So, matrix score is used to calculate 0 + 1*age + 2*smoking for all cases, and then summarise is used to 18 Lesson 7
derive the mean of this. The if newt == 1 qualifier is used to ensure the mean is taken over individual cases, rather than over time intervals. The following code illustrates the strategy as a whole:
. * First re-do earlier model . pgmhaz8 logt age smoking , id(id) seq(newt) dead(died) nolog < output omitted > . * Now generate predicted probabilities . mat b = e(b) . mat list b b[1,5] hazard: hazard: hazard: hazard: ln_varg: logt age smoking _cons _cons 2.7422156 .43690055 .95519812 -26.43864 .54098674
y1
. scalar n = colsof(b) - 1 . scalar list n n = . . mat b = b[1,2..n] . mat list b b[1,3] hazard: hazard: hazard: age smoking _cons .43690055 .95519812 -26.43864
4
/* exclude coeffs on dur dep var(s), and lnvarg est */
y1
. mat score xb = b . sum xb if newt == 1 Variable | Obs Mean Std. Dev. Min Max -------------+----------------------------------------------------xb | 80 -7.278995 4.992778 -15.07923 1.604393 . ge z0 = r(mean) + _b[logt]*logt . ge z1 = _b[_cons] + _b[age]*45 + _b[smoking]*1 + _b[logt]*logt . ge z2 = _b[_cons] + _b[age]*30 + _b[smoking]*0 + _b[logt]*logt . . sort id newt . by id: gen p0 = 1 - exp(-exp(z0)) . lab var p0 "Predicted h(t) at mean of covariates" . by id: gen p1 = 1 - exp(-exp(z1)) . lab var p1 "Predicted h(t),age=45,smoking" . by id: gen p2 = 1 - exp(-exp(z2)) . lab var p2 "Predicted h(t),age=30,non-smoking" . . by id: ge s0 =
exp(sum(ln(1-p0)))
. lab var s0 "Predicted S(t) at mean of covariates" . by id: ge s1 = exp(sum(ln(1-p1)))
. lab var s1 "Predicted S(t),age=45,smoking" . by id: ge s2 = exp(sum(ln(1-p2)))
19
Lesson 7
. lab var s2 "Predicted S(t),age=30,non-smoking"
Having generated the predicted hazard and survivor probabilities, we can graph them:
. . twoway (connect p0 newt , sort msymbol(t) ) /// > (connect p1 newt, sort msymbol(o) ) /// > (connect p2 newt, sort msymbol(x) ) /// > , title("Predicted discrete hazard rates from -pgmhaz8-") /// > saving(pgmh1, replace) ytitle("p(t)") (file pgmh1.gph saved)
. twoway (connect s0 newt , sort msymbol(t) ) /// > (connect s1 newt, sort msymbol(o) ) /// > (connect s2 newt , sort msymbol(x) ) /// > , title("Predicted survivor functions from -pgmhaz8-") /// > saving(pgms1, replace) ytitle("S(t)") (file pgms1.gph saved)
Figure 8.1. Predicted hazard rates from -pgmhaz8-
Predicted discrete hazard rates from -pgmhaz81 0 0 .2 .4 p(t) .6 .8
10
20 newt
30
40
Predicted h(t) at mean of covariates Predicted h(t),age=30,non-smoking
Predicted h(t),age=45,smokin
20
Lesson 7
Figure 8.2. Predicted survivor functions from -pgmhaz8-
Predicted survivor functions from -pgmhaz81 0 0 .2 .4 S(t) .6 .8
10
20 newt
30
40
Predicted S(t) at mean of covariates Predicted S(t),age=30,non-smoking
Predicted S(t),age=45,smokin
The median duration for the person with sample mean values is about 10, but clearly very much longer for the 30 year old non-smoker (out of sample range). The median duration for the 45 year-old smoker is shorter about six and the survival probability in this case is zero by the tenth interval. Observe from the first graph how the hazard rate for this person rises very quickly. Finally, we do a bit of tidying up:
. drop z0 z1 z2 xb p0 p1 p2 s0 s1 s2
This code only produces predictions for survival times ranging between the minimum and the maximum in the sample. If you want predictions for longer survival times, then you have to generate extra sample observations just as we did in Lesson 6. Observe that the same code could be used to generate predictions after estimating the proportional hazards or logistic models with Normal heterogeneity (the order in which the regressors are listed has to be the same as before for this to work without change). The commands would be: . xtcloglog
died logt age smoking , i(id) nolog nolog
. xtlogit died logt age smoking , i(id)
Note that, for these models, one could predict hazard rates in this model for each person within the sample using their built-in predict commands:
. predict p, pu0
Observe the pu0 option. This ensures that the prediction is generated while conditioning on frailty being set equal to its mean value. You could then generate the survivor function predictions with 21 Lesson 7
. bysort id (newt): ge s =
exp(sum(ln(1-p)))
The within-sample predictions could be listed and so on, as we did in Lesson 6.
5
(i)
Exercise 7.1
Repeat the estimation of frailty and non-frailty models using the bc.dta for a Log-logistic hazard model rather than a Weibull one. In what way do the results change? For any one of the three persons considered in the discrete-time model predictions, compare the predicted median survival times from the proportional hazards models with and without Normal heterogeneity (i.e. xtcloglog and cloglog with age and smoking as the regressors). Which model predicts the shorter median? Explain the result you find, referring to the parameter estimates from the two models. Repeat the estimation of Normal frailty and non-frailty models using the breast cancer data (bc.dta) for a discrete time logistic model rather than a discrete time proportional hazards model. Use logit and xtlogit rather than cloglog and xtcloglog. Compare model predictions too for the three persons considered in the text. Verify that uninformative episode-splitting does not affect the estimates of the continuous time frailty models. Use the Weibull model to do this and the following commands:
use bc, clear ge id = _n stset t, f(dead) id(id) streg age smoking , d(weib) nohr nolog frailty(gamma) stsplit time, every(1) stset streg age smoking , d(weib) nohr nolog frailty(gamma)
(ii)
(iii)
(iv)
(v)
Compare the results of the various frailty and non-frailty models, discrete and continuous, using a different data. I suggest the cancer data set (cancer.dta) rather than bc.dta. Using bc.dta, we compared the estimates derived using
pgmhaz8 logt age smoking , id(id) seq(newt) dead(died) nolog xtcloglog died logt age smoking , nolog i(id)
(vi)
These assume a parametric continuous distribution for the frailty mixture distribution. Re-estimate the model assuming a two mass point discrete mixture using the command
hshaz logt age smoking , id(id) seq(newt) dead(died) nolog
(By default, hshaz assumes that there are two mass points; the number can be changed using the nmp() option.) Comment on the estimates and compare them with those for the earlier models. Derive and compare the predicted hazard and survivor functions for persons from Type 1 and Type 2 (setting the other covariates at their mean values). 22 Lesson 7

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Learning Styles Exercise Please take a few moments to review the following illustration. You will be asked to use this illustration of cognitive and learning styles (CLS) to solve instructional design problems.holistic independent visualtextual visu

Wilfrid Laurier - GEOG - 415

1Geography 415 Hydrology LAB 3 (February 7, 2002)TA: Drew Lejbak Office: ES 310, 220-5590, atlejbak@ucalgary.ca Hours: Friday 10:00am 12:00pm, or by appointment Drop Box: Outside ES 313 1. Estimation of Net Radiation Table 1 lists the monthly mea

Wilfrid Laurier - GOPH - 457

GOPH457 - LAB 3 ELECTRICAL RESISTIVITY AND FORMATION FACTOR OBJECTIVE: To measure the electrical resistivity/conductivity of water saturated rock samples and to determine its formation factor. THEORY:L , where is the electrical A resistivity, L is

Wilfrid Laurier - ENSC - 501

1 ENSC 501 (Preliminary Draft) Environmental Science Field Course II Prepared by Ann-Lise Norman August 4, 2000 Coordinator: Dr. Ann-Lise Norman Principal Lecturers: Dr. Jamie Baxter (Geography) Dr. Angus Chu (Civil Engineering) Dr. Lis Dixon (Chemis

Wilfrid Laurier - ENCM - 369

University of Calgary Department of Electrical and Computer Engineering ENCM 369: Computer Organization Instructors: Dr. S. A. Norman (L01) and Dr. S. Yanushkevich (L02) Winter 2002 FINAL EXAMINATIONNote for Winter 2004 studentsIn order to save pa

Wilfrid Laurier - ENCM - 369

m02d08.txtTue Feb 08 10:07:52 20051ENCM 369 Tutorial for Feb. 8, 2005 NOTE ABOUT PUTTING LOCAL VARIABLES IN STACK SLOTS Some students doing Lab 3 Exercise D got confused about how to set up the local variables for main. foo and bar were ints, b

Wilfrid Laurier - ENCM - 369

ENCM 369 Tutorial for Feb. 7, 2006 NOTE ABOUT PUTTING LOCAL VARIABLES IN STACK SLOTS Some students doing Lab 3 Exercise D got confused about how to set up the local variables for main. local_a and local_b were ints, but their addresses were passed to

Wilfrid Laurier - ENCM - 369

University of Calgary Department of Electrical and Computer Engineering ENEL 369: Introduction to Computer Architecture Winter 2001 FINAL EXAMINATIONNote for Winter 2004 studentsIn order to save paper, I have reformatted this document to leave out

Brookdale - CS - 3172

Wilfrid Laurier - ENCM - 369

ENCM 369 Winter 2006: Lab 6, for the week of Feb. 27Author: Dr. S. A. Norman Electronic copies of handouts for ENCM 369 can be found by following links at http:/www.enel.ucalgary.ca/People/Norman/encm369winter2006/ Corrections to lab instructions wi

Wilfrid Laurier - ENGG - 233

ENGG 233 Fall 2004 L02: Examples of FUNCTION OVERLOADING and LIFETIME OF VARIABLES for Oct. 1Author: Dr. S. A. Norman Electronic copies of handouts for L02 can be found at http:/www.enel.ucalgary.ca/People/Norman/engg233-L02-fall2004/ func-overload.

Wilfrid Laurier - ENCM - 369

# ENCM 369 Winter 2002 Problem 8 Solution# S. Norman, April 30, 2002## Note that 2*31 can't be represented as a 32-bit int, but# - 2*31 can. The easiest way to deal with this is to do a# special check for - 2*31. After that we can ignore the#

Wilfrid Laurier - ENCM - 369

# SOLUTION to ENCM 369 Winter 2003 Final Exam Problem 2## NOTE: The cvt.d.w instruction was not listed in the 2003 Final# Exam Instruction Subset. During the exam, students were told# something like, "Problem 2 requires an instruction that isn't

Allan Hancock College - NWSAA - 20042

Western Australia Nuclear Waste Storage (Prohibition) Amendment Act 2004 Western Australia Nuclear Waste Storage (Prohibition) Amendment Act 2004 CONTENTS

Wilfrid Laurier - ENME - 337

enme_337_demo_2.nb1Example operations on lists (continued from "demo_1") Defining the array C (composed of known quantities) - this array is a list of 3 lists, each of them comprising 3 elementsCLHS i ry 0 j j j = j 0 ry j j j k 0 0 0 0 0 y z z

Wilfrid Laurier - ENME - 337

University of Calgary Department of Mechanical and Manufacturing EngineeringDecember 2, 2005Name: KeyCorrect answers are highlighted. Two versions of the Question Sheet contained the same questions in slightly changed order.ENME337 Computer T

Wilfrid Laurier - ENME - 337

116.820176.738-418.609121.196180.940-418.613116.820176.744-423.368121.212180.954-423.375

Air Force Academy - EDXSERVEG - 802

Air Force Academy - EDXSERVEG - 802

THE BACKBURNERTheres No Nobel Prize in EducationBy David WileyOn November 27, 1895, Alfred Nobel signed his last will in Paris, briefly outlining his vision of five prizes for those who during the past year have done humanity the greatest servic

Air Force Academy - STA - 575

September 9, 1994Letters to the EditorAAP News141 Northwest Point Blvd.P.O. Box 927Elk Grove Village, Ill. 60009-0927 Since 94% of "children's" firearm-related deaths1 are in anage group (10-19) constituting just 12% of all

Sveriges lantbruksuniversitet - E - 495

ENSC 495/851 LabENSC 495/851 Chip DesignIn the 495/851 lab you will build a 4 level device starting with an n <100 wafer, and doing 4 levels of definition. The mask levels are: Level 1 is a P (boron) diffusion to create pn diodes, p tubs (areas fo

Sveriges lantbruksuniversitet - E - 495

ENSC 495/851 Lab 2 WriteupStarted Feb. 25 2003 PurposeLab 2 continues on where lab 1 left off with levels 2 (n diffusion), 3 (contact cuts), 4 (metal) to produce the working 4 level device. With the n diffusion you have created the npn transistors,

Sveriges lantbruksuniversitet - E - 495

Vacuum Systems & Thin Films (Jaeger 6, Campbell 10&12, Ruska 7) Deposition of thin films involves vacuum system Direct Deposition from Source Evaporation Sputter Building up layers Chemical Vapour Deposition (CVD) Molecular Beam EpitaxyVacuum

Sveriges lantbruksuniversitet - E - 495

Diffusion and Ion implantation Reference: Chapter 4 Jaeger or Chapter 3 Ruska N & P Dopants determine the resistivity of material Note N lower resistavity than p: due to higher carrier mobility Near linear relationship below 0.2 ohm-cm (~1016 cm-3

Sveriges lantbruksuniversitet - E - 495

Etching Etch Definitions Isotropic Etching: same in all direction Anisotropic Etching: direction sensitive Selectivity: etch rate difference between 2 materials Other layers below one being etch Masking material (photoresist) Undercutting: materi

Sveriges lantbruksuniversitet - E - 495

Ion Implantation Most modern devices doped using ion implanters Ionize gas sources (single +, 2+ or 3+ ionization) Accelerate dopant ions to very high voltages (10-600 KeV) Use analyzer to selection charge/mass ratio (ie ionization) Bend beam to

Air Force Academy - STA - 575

NEWS RELEASENovember 27, 1997 For Immediate ReleaseSPEAKER'S RULINGS DENY M.P.s' RIGHTS FOR 4TH TIME IN A ROW"Political parties should not control M.P.s' rights and privileges inParliament."Ottawa - Today, Garry Breitkreuz, M.P. for

Air Force Academy - STA - 575

From: owner-cdn-firearms-digest@sfn.saskatoon.sk.ca on behalf ofCdn-Firearms Digest [owner-cdn-firearms-digest@sfn.saskatoon.sk.ca]Sent: Friday, 21 December, 2001 11:34To: cdn-firearms-digest@broadway.sfn.saskatoon.sk.caSubject: Cdn-Firearms Dige

Sveriges lantbruksuniversitet - ENSC - 460

University of Toronto ECE532 Digital Hardware Module m05: Adding a User-Designed Peripheral Version for EDK 8.2.02i as of January 7, 2007New in EDK 7.1iThe Create and Import Peripheral Wizard was introduced in EDK 7.1i to simplify adding user-desi

Sveriges lantbruksuniversitet - ENSC - 460

Figure 1-1 System Block DiagramVGA monitorKeyboardVGA Interface (DAC)Multimedia BoardGraphics bridge/ controller (custom)Input Interface (custom)Xilinx Virtex-II FPGAMB Debug module (Xilinx IP) Microblaze Processor OPB Interface (Xili