hip <- read.table("http://astrostatistics.psu.edu/datasets/HIP_star.dat", header=T,fill=T) attach(hip) filter1 <- (RA>50 & RA<100 & DE>0 & DE<25) filter2 <- (pmRA>90 & pmRA<130 & pmDE>-60 & pmDE< -10) filter <- filter1 & filter2 & (e_Plx<5) sum(filter)In this section of the tutorial, we will compare these Hyades stars with the remaining stars in the Hipparcos dataset on the basis of the color (B minus V) variable. That is, we are comparing the groups in the boxplot below:

color <- B.V boxplot(color~filter,notch=T)For ease of notation, we define vectors H and nH (for "Hyades" and "not Hyades") that contain the data values for the two groups.

H <- color[filter] nH <- color[!filter & !is.na(color)]In the definition of nH above, we needed to exclude the NA values.

A two-sample t-test may now be performed with a single line:

t.test(H,nH)Because it is instructive and quite easy, we may obtain the same results without resorting to the t.test function. First, we calculate the variances of the sample means for each group:

v1 <- var(H)/92 v2 <- var(nH)/2586 c(var(H),var(nH))The t statistic is based on the standardized difference between the two sample means. Because the two samples are assumed independent, the variance of this difference equals the sum of the individual variances (i.e., v1+v2). Nearly always in a two-sample t-test, we wish to test the null hypothesis that the true difference in means equals zero. Thus, standardizing the difference in means involves subtracting zero and then dividing by the square root of the variance:

tstat <- (mean(H)-mean(nH))/sqrt(v1+v2) tstatTo test the null hypothesis, this t statistic is compared to a t distribution. In a Welch test, we assume that the variances of the two populations are not necessarily equal, and the degrees of freedom of the t distribution are computed using the so-called Satterthwaite approximation:

(v1 + v2)^2 / (v1^2/91 + v2^2/2585)The two-sided p-value may now be determined by using the cumulative distribution function of the t distribution, which is given by the pt function.

2*pt(tstat,97.534)Incidentally, one of the assumptions of the t-test, namely that each of the two underlying populations is normally distributed, is almost certainly not true in this example. However, because of the central limit theorem, the t-test is robust against violations of this assumption; even if the populations are not roughly normally distributed, the sample means are.

In this particular example, the Welch test is probably not necessary, since the sample variances are so close that an assumption of equal variances is warranted. Thus, we might conduct a slightly more restrictive t-test that assumes equal population variances. Without going into the details here, we merely present the R output:

t.test(H,nH,var.equal=T)

gc <- read.csv("http://astrostatistics.psu.edu/datasets/glob_clus.csv") t.test(lum~gal, data=gc, mu=24.44)As before, the t-statistic here (1.6254) is accompanied by a p-value (0.1074). This p-value may be interpreted as follows: IF the two samples really are representive samples from populations whose means differ by 24.44, THEN the probability of obtaining a t-statistic at least as far from zero as 1.6254 would be roughly 0.1074. Since most people don't consider 0.1074 to be a very small probability, the conclusion here is that there is not strong statistical evidence of a difference in true means other than 24.44.

The p-value of 0.1074 is an approximation calculated under certain distributional assumptions on the populations. An alternative method for calculating a p-value corresponding to the t-statistic of 1.6254 is called a permutation test. Conceptually, the permutation test follows directly from the definition (given in the preceding paragraph) of a p-value. Let us assume that the luminosities for the MWG globular clusters are distributed exactly like the luminosities for the M31 globular clusters, except that the latter are all shifted down by 24.44. Under this assumption (called the

By definition, then, the p-value should be equal to the proportion of all possible t-statistics resulting from label permutations that are at least as extreme as the observed t-statistic (1.6254). Since there are far too many such permuations (roughly 10

tlist <- 1:5000 lum <- gc[,"lum"] gal <- gc[,"gal"] lum[gal=="M31"] <- lum[gal=="M31"]-24.44 for(i in 1:5000) { s <- sample(441,81) # choose a sample newtstat <- t.test(lum[s], lum[-s])$stat tlist[i] <- newtstat # add new null t-stat to list }Note: The above code is

By definition, the p-value is the probability of obtaining a test statistic more extreme than the observed test statistic under the null hypothesis. Let's take a look at the null distribution of the t statistic we just calculated, along with the observed value:

hist(tlist) abline(v=c(-1,1)*1.6254,lty=2,col=2) sum(abs(tlist)>=1.6254)/5000This p-value is quite close to the 0.1074 obtained earlier.

A simple yet very powerful graphical device is called a Q-Q plot, in which some quantiles of the sample are plotted against the same quantiles of whatever distribution we have in mind. If a roughly straight line results, this suggests that the fit is good. Roughly, a pth quantile of a distribution is a value such that a proportion p of the distribution lies below that value.

A Q-Q plot for normality is so common that there is a separate function, qqnorm, that implements it. (Normality is vital in statistics not merely because many common populations are normally distributed -- which is actually not true in astronomy -- but because the central limit theorem guarantees the approximate normality of sample means.)

par(mfrow=c(2,2)) qqnorm(tlist,main="Null luminosity t statistics") abline(0,1,col=2) qqnorm(H,main="Hyades") qqnorm(nH,main="non-Hyades")Not surprisingly, the tlist variable appears extremely nearly normally distributed (more precisely, it is nearly

In the bottom right corner of the plotting window, let's reconstruct the Q-Q plot from scratch for tlist. This is instructive because the same technique may be applied to any comparison distribution, not just normal. If we consider the 5000 entries of tlist in increasing order, let's call the ith value the ((2i-1)/10000)th quantile for all i from 1 to 5000. We merely graph this point against the corresponding quantile of standard normal:

plot(qnorm((2*(1:5000)-1)/10000),sort(tlist)) par(mfrow=c(1,1)) # reset plotting windowRelated to the Q-Q plot is a distribution function called the

plot(ecdf(tlist))While it is generally very difficult to interpret the EDF directly, it is possible to compare an EDF to a theoretical cumulative distribution function or two another EDF. Among the statistical tests that implement such a comparison is the Kolmogorov-Smirnov test, which is implemented by the R function ks.test.

ks.test(tlist,"pnorm") ks.test(H,nH)Whereas the first result above gives a surprisingly small p-value, the second result is not surprising; we already saw that H and nH have statistically significantly different means. However, if we center each, we obtain

ks.test(H-mean(H),nH-mean(nH))In other words, the Kolmogorov-Smirnov test finds no statistically significant evidence that the distribution of B.V for the Hyades stars is anything other than a shifted version of the distribution of B.V for the other stars. We can perform the same test for the luminosities of globular clusters from both the Milky Way and Andromeda:

lum.mwg <- gc[gal=="MWG","lum"] lum.m31 <- gc[gal=="M31","lum"] ks.test(lum.mwg, lum.m31-24.44)The above test does give a statistically significant difference. The K-S test tests whether the M31 dataset, shifted by 24.44, comes from the same distribution as the MWG dataset. The t-test, on the other hand, only tests whether these distributions have the same mean.

bvcat <- cut(color, breaks=c(-Inf,.5,.75,1,Inf)) boxplot(Vmag~bvcat, varwidth=T, ylim=c(max(Vmag),min(Vmag)), xlab=expression("B minus V"), ylab=expression("V magnitude"), cex.lab=1.4, cex.axis=.8)The cut values for bvcat are based roughly on the quartiles of the B minus V variable. We have created, albeit artificially, a second categorical variable ("filter", the Hyades indicator, is the first). Here is a summary of the dataset based only on these two variables:

table(bvcat,filter)Note that the Vmag variable is irrelevant in the table above.

To perform a chi-squared test of the null hypothesis that the true population proportions falling in the four categories are the same for both the Hyades and non-Hyades stars, use the chisq.test function:

chisq.test(bvcat,filter)Since we already know these two groups differ with respect to the B.V variable, the result of this test is not too surprising. But it does give a qualitatively different way to compare these two distributions than simply comparing their means.

The p-value produced above is based on the fact that the chi-squared statistic is approximately distributed like a true chi-squared distribution (on 3 degrees of freedom, in this case) if the null hypothesis is true. However, it is possible to obtain exact p-values, if one wishes to calculate the chi-squared statistic for all possible tables of counts with the same row and column sums as the given table. Since this is rarely practical computationally, the exact p-value may be approximated using a Monte Carlo method (just as we did earlier for the permutation test). Such a method is implemented in the chisq.test function:

chisq.test(bvcat,filter,sim=T,B=50000)The two different p-values we just generated a numerically similar but based on entirely different mathematics. The difference may be summed up as follows: The first method produces the exact value of an approximate p-value, whereas the second method produces an approximation to the exact p-value!

The test above is usually called a chi-squared test of homogeneity. If we observe only one sample, but we wish to test whether the categories occur in some pre-specified proportions, a similar test (and the same R function) may be applied. In this case, the test is usually called a chi-squared test of goodness-of-fit.

x <- DE[filter] y <- pmDE[filter] plot(x,y,pch=20) model1 <- lm(y ~ x) abline(model1,lwd=2,col=2)The red line on the plot is the usual least-squares line, for which estimation is easy and asymptotic theory gives easy-to-calculate standard errors for the coefficients:

summary(model1)$coefHowever, suppose we wish to use a resistant regression method such as lqs.

library(MASS) model2 <- lqs(y ~ x) abline(model2,lwd=2,col=3) model2In this case, it is not so easy to obtain standard errors for the coefficients. Thus, we will turn to bootstrapping. In a standard, or nonparametric, bootstrap, we repeatedly draw samples of size 92 from the empirical distribution of the data, which in this case consist of the (DE, pmDE) pairs. We use lqs to fit a line to each sample, then compute the sample covariance of the resulting coefficient vectors. The procedure works like this:

model2B <- matrix(0,200,2) for (i in 1:200) { s <- sample(92,replace=T) model2B[i,] <- lqs(y[s]~x[s])$coef }We may now find the sample covariance matrix for model2B. The (marginal) standard errors of the coefficients are obtained as the square roots of the diagonal entries of this matrix:

cov(model2B) se <- sqrt(diag(cov(model2B))) seThe logic of the bootstrap procedure is that we are estimating an approximation of the true standard errors. The approximation involves replacing the true distribution of the data (unknown) with the empirical distribution of the data. This approximation may be estimated with arbitrary accuracy by a Monte Carlo approach, since the empirical distribution of the data is known and in principle we may sample from it as many times as we wish. In other words, as the bootstrap sample size increases, we get a better estimate of the true value of the

An alternative way to generate a bootstrap sample in this example is by generating a new value of each response variable (y) by adding the predicted value from the original lqs model to a randomly selected residual from the original set of residuals. Thus, we resample not the entire bivariate structure but merely the residuals. As an exercise, you might try implementing this approach in R. Note that this approach is not a good idea if you have reason to believe that the distribution of residuals is not the same for all points. For instance, if there is heteroscedasticity or if the residuals contain structure not explained by the model, this residual resampling approach is not warranted.

library(boot) mystat <- function(a,b) lqs(a[b,2]~a[b,1])$coef model2B.2 <- boot(cbind(x,y), mystat, 200) names(model2B.2)As explained in the help file, the boot function requires as input a function that accepts as arguments the whole dataset and an index that references an observation from that dataset. This is why we defined the mystat function above. To see the output that is similar to that obtained earlier for the m2B object, look in m2B2$t:

cov(model2B.2$t) sqrt(diag(cov(model2B.2$t)))Compare with the output provided by print.boot and the plot produced by plot.boot:

model2B.2 plot(model2B.2)Another related function, for producing bootstrap confidence intervals, is boot.ci.

Sometimes, resampling is done from a theoretical distribution rather than from the original sample. For instance, if simple linear regression is applied to the regression of pmDE on DE, we obtain a parametric estimate of the distribution of the residuals, namely, normal with mean zero and standard deviation estimated from the regression:

summary(model1)Remember that model1 was defined above as lm(y~x). We observe a residual standard error of 4.449.

A parametric bootstrap scheme proceeds by simulating a new set of pmDE (or y) values using the model

y <- 21.9 - 3.007*x + 4.449*rnorm(92)Then, we refit a linear model using y as the new response, obtaining slightly different values of the regression coefficients. If this is repeated, we obtain an approximation of the joint distribution of the regression coefficients for this model.

Naturally, the same approach could be tried with other regression methods such as those discussed in the EDA and regression tutorial, but careful thought should be given to the parametric model used to generate the new residuals. In the normal case discussed here, the parametric bootstrap is simple, but it is really not necessary because standard linear regression already gives a very good approximation to the joint distribution of the regression coefficients when errors are heteroscedastic and normal. One possible use of this method is in a model that assumes the absolute residuals are exponentially distributed, in which case least absolute deviation regression as discussed in the EDA and regression tutorial can be justified. The reader is encouraged to implement a parametric bootstrap using the rq function found in the "quantreg" package.

tlist2 <- NULL all <- c(H,nH) for(i in 1:5000) { s <- sample(2586,92) # choose a sample tlist2 <- c(tlist2, t.test(all[s],all[-s], var.eq=T)$stat) # add t-stat to list }Let's look at two different ways of assessing whether the values in the tlist2 vector appear to be a random sample from a standard normal distribution. The first is graphical, and the second uses the Kolmogorov-Smirnov test.

plot(qnorm((2*(1:5000)-1)/10000), sort(tlist2)) abline(0,1,col=2) ks.test(tlist2, "pnorm")The graphical method does not show any major deviation from standard normality, but this graphical "test" is better able to detect departures from an overall normal shape than to detect, say, a shift of the mean away from zero. The following procedures illustrate this fact:

plot(qnorm((2*(1:5000)-1)/10000), sort(tlist2)-mean(tlist2)) abline(0,1,col=2) ks.test(tlist2-mean(tlist2), "pnorm")The graphical plot shows no discernable difference from before, but we see a vast difference in the new p-value returned by the Kolmogorov-Smirnov test (!!).

However, let us now consider the p-value returned by the last use of ks.test above. It is not quite valid because the theoretical null distribution against which we are testing depends upon an estimate (the mean) derived from the data. To get a more accurate p-value, we may use a bootstrap approach.

First, obtain the Kolmogorov-Smirnov test statistic from the test above:

obs.ksstat <- ks.test(tlist2-mean(tlist2), "pnorm")$statNow we'll generate a new bunch of these statistics under the null hypothesis that tlist2 really represents a random sample from

random.ksstat <- NULL for(i in 1:1000) { x <- rnorm(5000) random.ksstat <- c(random.ksstat, ks.test(x,pnorm,mean=mean(x))$stat) }Now let's look at a histogram of the test statistics and estimate a p-value:

hist(random.ksstat,nclass=40) abline(v=obs.ksstat,lty=2,col=2) mean(random.ksstat>=obs.ksstat)Note that the approximate p-value above is smaller than the original p-value reported by newkstest, though it is still not small enough to provide strong evidence that the tlist2 sample is not normal with unit variance.

The bootstrap procedure above relied on multiple resamples with replacement. Since these samples were drawn from a theoretical population (in this case, a normal distribution with parameters that might be determined by the data), it is considered a

In a

s <- sample(5000,replace=T) ks.test(tlist[s],pnorm,mean=mean(tlist[s]))$statHowever, this statistic has a bias that must be corrected. To make this correction, it is necessary to rewrite the ks.test function itself. There is no way to use the output of the ks.test function to produce the bias-corrected statistics.