Hypothesis Testing COMP 245 STATISTICS. Dr N A Heard. 1 Hypothesis Testing Introduction Error Rates and Power of a Test...


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1 Hypothesis Testing COMP 45 STATISTICS Dr N A Heard Contents 1 Hypothesis Testing 1.1 Introduction Error Rates and Power of a Test Testing for a population mean 3.1 Normal Distribution with Known Variance Duality with Confidence Intervals Normal Distribution with Unknown Variance Testing for differences in population means Two Sample Problems Normal Distributions with Known Variances Normal Distributions with Unknown Variances Goodness of Fit Count Data and ChiSquare Tests Proportions Model Checking Independence
2 1 Hypothesis Testing 1.1 Introduction Hypotheses Suppose we are going to obtain a random i.i.d. sample X = (X 1,..., X n ) of a random variable X with an unknown distribution P X. To proceed with modelling the underlying population, we might hypothesise probability models for P X and then test whether such hypotheses seem plausible in light of the realised data x = (x 1,..., x n ). Or, more specifically, we might fix upon a parametric family P X θ with unknown parameter θ and then hypothesise values for θ. Generically let θ 0 denote a hypothesised value for θ. Then after observing the data, we wish to test whether we can indeed reasonably assume θ = θ 0. For example, if X N(µ, σ ) we may wish to test whether µ = 0 is plausible in light of the data x. Formally, we define a null hypothesis H 0 as our hypothesised model of interest, and also specify an alternative hypothesis H 1 of rival models against which we wish to test H 0. Most often we simply test H 0 : θ = θ 0 against H 1 : θ = θ 0. This is known as a twosided test. In some situations it may be more appropriate to consider alternatives of the form H 1 : θ > θ 0 or H 1 : θ < θ 0, known as onesided tests. Rejection Region for a Test Statistic To test the validity of H 0, we first choose a test statistic T(X) of the data for which we can find the distribution, P T, under H 0. Then, we identify a rejection region R R of low probability values of T under the assumption that H 0 is true, so P(T R H 0 ) = α for some small probability α (typically 5%). A well chosen rejection region will have relatively high probability under H 1, whilst retaining low probability under H 0. Finally, we calculate the observed test statistic t(x) for our observed data x. If t R we reject the null hypothesis at the 100α% level. pvalues For each possible significance level α (0, 1), a hypothesis test at the 100α% level will result in either rejecting or not rejecting H 0. As α 0 it becomes less and less likely that the null hypothesis will be rejected, as the rejection region is becoming smaller and smaller. Similarly, as α 1 it becomes more and more likely that the null hypothesis will be rejected. For a given data set and resulting test statistic, we might, therefore, be interested in identifying the critical significance level which marks the threshold between us rejecting and not rejecting the null hypothesis. This is known as the pvalue of the data. Smaller pvalues suggest stronger evidence against H Error Rates and Power of a Test Test Errors There are two types of error in the outcome of a hypothesis test:
3 Type I: Rejecting H 0 when in fact H 0 is true. By construction, this happens with probability α. For this reason, the significance level of a hypothesis test is also referred to as the Type I error rate. Type II: Not rejecting H 0 when in fact H 1 is true. The probability with which this type of error will occur depends on the unknown true value of θ, so to calculate values we plugin a plausible alternative value for θ = θ 0, θ 1 say, and let β = P(T / R θ = θ 1 ) be the probability of a Type II error. Power We define the power of a hypothesis test by 1 β = P(T R θ = θ 1 ). For a fixed significance level α, a well chosen test statistic T and rejection region R will have high power  that is, maximise the probability of rejecting the null hypothesis when the alternative is true. Testing for a population mean.1 Normal Distribution with Known Variance N(µ, σ )  σ Known Suppose X 1,..., X n are i.i.d. N(µ, σ ) with σ known and µ unknown. We may wish to test if µ = µ 0 for some specific value µ 0 (e.g. µ 0 = 0, µ 0 = 9.8). Then we can state our null and alternative hypotheses as H 0 :µ = µ 0 ; H 1 :µ = µ 0. Under H 0 : µ = µ 0, we then know both µ and σ. So for the sample mean X we have a known distribution for the test statistic Z = X µ 0 σ/ n Φ. So if we define our rejection region R to be the 100α% tails of the standard normal distribution distribution, ) ( ) R = (, z 1 α z 1 α, { } z z > z 1 α, we have P(Z R) = α under H 0. We thus reject H 0 at the 100α% significance level our observed test statistic z = x µ 0 σ/ n R. The pvalue is given by {1 Φ( z )}. 3
4 Ex. 5% Rejection Region for N(0, 1) Statistic φ(z) R z.1.1 Duality with Confidence Intervals There is a strong connection in this context between hypothesis testing and confidence intervals. Suppose we have constructed a 100(1 α)% confidence interval for µ. Then this is precisely the set of values {µ 0 } for which there would be insufficient evidence to reject a null hypothesis H 0 : µ = µ 0 at the 100α%level. Example A company makes packets of snack foods. The bags are labelled as weighing 454g; of course they won t all be exactly 454g, and let s suppose the variance of bag weights is known to be 70g². The following data show the mass in grams of 50 randomly sampled packets. 464, 450, 450, 456, 45, 433, 446, 446, 450, 447, 44, 438, 45, 447, 460, 450, 453, 456, 446, 433, 448, 450, 439, 45, 459, 454, 456, 454, 45, 449, 463, 449, 447, 466, 446, 447, 450, 449, 457, 464, 468, 447, 433, 464, 469, 457, 454, 451, 453, 443 Are these data consistent with the claim that the mean weight of packets is 454g? 1. We wish to test H 0 : µ = 454 vs. H 1 : µ = 454. So set µ 0 = Although we have not been told that the packet weights are individually normally distributed, by the CLT we still have that the mean weight of the sample of packets is approximately normally distributed, and hence we still approximately have Z = X µ 0 σ/ n Φ. 3. x = 451. and n = 50 z = x µ 0 σ/ n = For a 5%level significance test, we compare the statistic z =.350 with the rejection region R = (, z ) (z 0.975, ) = (, 1.96) (1.96, ). Clearly we have z R, and so at the 5%level we reject the null hypothesis that the mean packet weight is 454g. 4
5 5. At which significance levels would we have not rejected the null hypothesis? For a 1%level significance test, the rejection region would have been R = (, z ) (z 0.995, ) = (,.576) (.576, ). In which case z / R, and so at the 1%level we would not have rejected the null hypothesis. The pvalue is {1 Φ( z )} = {1 Φ(.350 )} ( ) = 0.019, and so we would only reject the null hypothesis for α > 1.9%.. Normal Distribution with Unknown Variance N(µ, σ )  σ Unknown Similarly, if σ in the above example were unknown, we still have that T = X µ 0 s n 1 / n t n 1. So for a test of H 0 : µ = µ 0 vs. H 1 : µ = µ 0 at the α level, the rejection region of our observed test statistic t = x µ 0 s n 1 / n is ) ( ) R = (, t n 1,1 α t n 1,1 α, { } t t > t n 1,1 α. Ex. 5% Rejection Region for t 5 Statistic f(t) R t 5
6 Example 1 Consider again the snack food weights example. There, we assumed the variance of bag weights was known to be 70. Without this, we could have estimated the variance by s n 1 = 1 n 1 n i=1 Then the corresponding tstatistic becomes very similar to the zstatistic of before. (x i x) = t = x µ 0 s n 1 / n =.341, And since n = 50, we compare with the t 49 distribution which is approximately N(0, 1). So the hypothesis test results and pvalue would be practically identical. Example A particular piece of code takes a random time to run on a computer, but the average time is known to be 6 seconds. The programmer tries an alternative optimisation in compilation and wishes to know whether the mean run time has changed. To explore this, he runs the reoptimised code 16 times, obtaining a sample mean run time of 5.8 seconds and biascorrected sample standard deviation of 1. seconds. Is the code any faster? 1. We wish to test H 0 : µ = 6 vs. H 1 : µ = 6. So set µ 0 = 6.. Assuming the run times are approximately normal, T = X µ 0 s n 1 / n t n 1. That is, X 6 s n 1 / 16 t 15. So we reject H 0 at the 100α% level if T > t 15,1 α/. 3. x = 5.8, s n 1 = 1. and n = 16 t = x µ 0 s n 1 / n = We have t = /3 <<.13 = t 15,.975, so we have insufficient evidence to reject H 0 at the 5% level. 5. In fact, the pvalue for these data is 51.51%, so there is very little evidence to suggest the code is now any faster. 6
7 3 Testing for differences in population means 3.1 Two Sample Problems Samples from Populations Suppose, as before, we have a random sample X = (X 1,..., X n1 ) from an unknown population distribution P X. But now, suppose we have a further random sample Y = (Y 1,..., Y n ) from a second, different population P Y. Then we may wish to test hypotheses concerning the similarity of the two distributions P X and P Y. In particular, we are often interested in testing whether P X and P Y have equal means. That is, to test H 0 : µ X = µ Y vs H 1 : µ X = µ Y. Paired Data A special case is when the two samples X and Y are paired. That is, if n 1 = n = n and the data are collected as pairs (X 1, Y 1 ),..., (X n, Y n ) so that, for each i, X i and Y i are possibly dependent. For example, we might have a random sample of n individuals and X i represents the heart rate of the i th person before light exercise and Y i the heart rate of the same person afterwards. In this special case, for a test of equal means we can consider the sample of differences Z 1 = X 1 Y 1,..., Z n = X n Y n and test H 0 : µ Z = 0 using the single sample methods we have seen. In the above example, this would test whether light exercise causes a change in heart rate. 3. Normal Distributions with Known Variances N(µ X, σ X ), N(µ Y, σ Y ) Suppose X = (X 1,..., X n1 ) are i.i.d. N(µ X, σ X ) with µ X unknown; Y = (Y 1,..., Y n ) are i.i.d. N(µ Y, σ Y ) with µ Y unknown; the two samples X and Y are independent. Then we still have that, independently, ( ) X N µ X, σ X n 1 ( ) Ȳ N µ Y, σ Y n From this it follows that the difference in sample means, ( ) X Ȳ N µ X µ Y, σ X + σ Y, n 1 n 7
8 and hence ( X Ȳ) (µ X µ Y ) σ X /n 1 + σ Y /n Φ. So under the null hypothesis H 0 : µ X = µ Y, we have Z = X Ȳ Φ. σx /n 1 + σy /n N(µ X, σx ), N(µ Y, σy )  σ X, σ Y Known So if σx and σ Y are known, we immediately have a test statistic z = x ȳ σx /n 1 + σy /n which we can compare against the quantiles of a standard normal. That is, { } R = z z > z 1 α, gives a rejection region for a hypothesis test of H 0 : µ X = µ Y vs. H 1 : µ X = µ Y at the 100α% level. 3.3 Normal Distributions with Unknown Variances N(µ X, σ ), N(µ Y, σ )  σ Unknown On the other hand, suppose σx and σ Y are unknown. Then if we know σ X = σ Y = σ but σ is unknown, we can still proceed. We have and so, under H 0 : µ X = µ Y, but with σ unknown. ( X Ȳ) (µ X µ Y ) σ 1/n 1 + 1/n Φ, X Ȳ σ 1/n 1 + 1/n Φ. Pooled Estimate of Population Variance We need an estimator for the variance using samples from two populations with different means. Just combining the samples together into one big sample would overestimate the variance, since some of the variability in the samples would be due to the difference in µ X and µ Y. So we define the biascorrected pooled sample variance Sn 1 +n = n 1 i=1 (X i X) + n i=1 (Y i Ȳ), n 1 + n 8
9 which is an unbiased estimator for σ. We can immediately see that s n 1 +n is indeed an unbiased estimate of σ by noting S n 1 +n = n 1 1 n 1 + n S n n 1 n 1 + n S n 1 ; That is, s n 1 +n is a weighted average of the biascorrected sample variances for the individual samples x and y, which are both unbiased estimates for σ. Then substituting S n1 +n in for σ we get and so, under H 0 : µ X = µ Y, ( X Ȳ) (µ X µ Y ) t n1 +n 1/n1 + 1/n, S n1 +n T = S n1 +n X Ȳ t n1 +n 1/n1 + 1/n. So we have a rejection region for a hypothesis test of H 0 : µ X = µ Y vs. H 1 : µ X = µ Y at the 100α% level given by for the statistic { R = t t > t n1 +n,1 α }, t = s n1 +n x ȳ. 1/n1 + 1/n Example The same piece of C code was repeatedly run after compilation under two different C compilers, and the run times under each compiler were recorded. The sample mean and biascorrected sample standard deviation for Compiler 1 were 114s and 310s respectively, and the corresponding figures for Compiler were 94s and 90s. Both sets of data were each based on 15 runs. Suppose that Compiler is a refined version of Compiler 1, and so if µ 1, µ are the expected run times of the code under the two compilations, we might fairly assume µ µ 1. Conduct a hypothesis test of H 0 : µ 1 = µ vs H 1 : µ 1 > µ at the 5% level. Until now we have mostly considered twosided tests. That is tests of the form H 0 : θ = θ 0 vs H 1 : θ = θ 0. Here we need to consider onesided tests, which differ by the alternative hypothesis being of the form H 1 : θ < θ 0 or H 1 : θ > θ 0. This presents no extra methodological challenge and requires only a slight adjustment in the construction of the rejection region. We still use the tstatistic t = s n1 +n x ȳ, 1/n1 + 1/n 9
10 where x, ȳ are the sample mean run times under Compilers 1 and respectively. But now the onesided rejection region becomes R = {t t > t n1 +n,1 α }. First calculating the biascorrected pooled sample variance, we get s n 1 +n = = 300. (Note that since the sample sizes n 1 and n are equal, the pooled estimate of the variance is the average of the individual estimates.) So t = s n1 +n = 10 = x ȳ = 1/n1 + 1/n /15 + 1/15 For a 1sided test we compare t = 3.16 with t 8,0.95 = and conclude that we reject the null hypothesis at the 5% level; the second compilation is significantly faster. 10
11 4 Goodness of Fit 4.1 Count Data and ChiSquare Tests Count Data The results in the previous sections relied upon the data being either normally distributed, or at least through the CLT having the sample mean being approximately normally distributed. Tests were then developed for making inference on population means under those assumptions. These tests were very much modelbased. Another important but very different problem concerns model checking, which can be addressed through a more general consideration of count data for simple (discrete and finite) distributions. The following ideas can then be trivially extended to infinite range discrete and continuous r.v.s by binning observed samples into a finite collection of predefined intervals. Samples from a Simple Random Variable Let X be a simple random variable taking values in the range {x 1,..., x k }, with probability mass function p j = P(X = x j ), j = 1,..., k. A random sample of size n from the distribution of X can be summarised by the observed frequency counts O = (O 1,..., O k ) at the points x 1,..., x k (so k j=1 O j = n). Suppose it is hypothesised that the true pmf {p j } is from a particular parametric model p j = P(X = x j θ), j = 1,..., k for some unknown parameter pvector θ. To test this hypothesis about the model, we first need to estimate the unknown parameters θ so that we are able to calculate the distribution of any statistic under the null hypothesis H 0 : p j = P(X = x j θ), j = 1,..., k. Let ˆθ be such an estimator, obtained using the sample O. Then under H 0 we have estimated probabilities for the pmf ˆp j = P(X = x j ˆθ), j = 1,..., k and so we are able to calculate estimated expected frequency counts E = (E 1,..., E k ) by E j = n ˆp j. (Note again we have k j=1 E j = n.) We then seek to compare the observed frequencies with the expected frequencies to test for goodness of fit. ChiSquare Test To test H 0 : p j = P(X = x j θ) vs. H 1 : 0 p j 1, p j = 1 we use the chisquare statistic X = k (O i E i ). E i=1 i If H 0 were true, then the statistic X would approximately follow a chisquare distribution with ν = k p 1 degrees of freedom. k is the number of values (categories) the simple r.v. X can take. p is the number of parameters being estimated (dim(θ)). For the approximation to be valid, we should have j, E j 5. This may require some merging of categories. 11
12 χ ν pdf for ν = 1,, 3, 5, 10 χ ν(x) χ 1 χ χ 3 χ 5 χ x Rejection Region Clearly larger values of X correspond to larger deviations from the null hypothesis model. That is, if X = 0 the observed counts exactly match those expected under H 0. For this reason, we always perform a onesided goodness of fit test using the χ statistic, looking only at the upper tail of the distribution. Hence the rejection region for a goodness of fit hypothesis test at the at the 100α% level is given by R = { } x x > χ k p 1,1 α. 4. Proportions Example Each year, around 1.3 million people in the USA suffer adverse drug effects (ADEs). A study in the Journal of the American Medical Association (July 5, 1995) gave the causes of 95 ADEs below. Cause Number of ADEs Lack of knowledge of drug 9 Rule violation 17 Faulty dose checking 13 Slips 9 Other 7 Test whether the true percentages of ADEs differ across the 5 causes. Under the null hypothesis that the 5 causes are equally likely, we would have expected counts of 95 5 = 19 for each cause. 1
13 So our χ statistic becomes x = (9 19) 19 + (17 19) 19 + (13 19) 19 = = = (9 19) 19 + (7 19) 19 We have not estimated any parameters from the data, so we compare x with the quantiles of the χ 5 1 = χ 4 distribution. Well 16 > 9.49 = χ 4,0.95, so we reject the null hypothesis at the 5% level; we have reason to suppose that there is a difference in the true percentages across the different causes. 4.3 Model Checking Example  Fitting a Poisson Distribution to Data Recall the example from the Discrete Random Variables chapter, where the number of particles emitted by a radioactive substance which reached a Geiger counter was measured for 608 time intervals, each of length 7.5 seconds. We fitted a Poisson(λ) distribution to the data by plugging in the sample mean number of counts (3.870) for the rate parameter λ. (Which we now know to be the MLE!) x O(n x ) E(n x ) (O=Observed, E=Expected). Whilst the fitted Poisson(3.87) expected frequencies looked quite convincing to the eye, at that time we had no formal method of quantitatively assessing the fit. However, we now know how to proceed. x O E O E (O E) E The statistic x = (O E) E = should be compared with a χ = χ 9 distribution. Well χ 9,0.95 = 16.91, so at the 5% level we do not reject the null hypothesis of a Poisson(3.87) model for the data. 4.4 Independence Contingency Tables Suppose we have two simple random variables X and Y which are jointly distributed with unknown probability mass function p XY. We are often interested in trying to ascertain whether X and Y are independent. That is, determine whether p XY (x, y) = p X (x)p Y (y). 13
14 Let the ranges of the r.v.s X and Y be {x 1,..., x k } and {y 1,..., y l } respectively. Then an i.i.d. sample of size n from the joint distribution of (X, Y) can be represented by a list of counts n ij (1 i k; 1 j l) of the number of times we observe the pair (x i, y j ). Tabulating these data in the following way gives what is known as a k l contingency table. y 1 y... y l x 1 n 11 n 1 n 1l n 1 x n 1 n n l n. x k n k1 n k n kl n k n 1 n... n l n Note the row sums (n 1, n,..., n k ) represent the frequencies of x 1, x,..., x k in the sample (that is, ignoring the value of Y). Similarly for the column sums (n 1, n,..., n l ) and y 1,..., y l. Under the null hypothesis H 0 : X and Y are independent, the expected values of the entries of the contingency table, conditional on the row and column sums, can be estimated by ˆn ij = n i n j, 1 i k, 1 j l. n To see this, consider the marginal distribution of X; we could estimate p X (x i ) by ˆp i = n i n. Similarly for p Y (y j ) we get ˆp j = n j n. Then under the null hypothesis of independence p XY (x i, y j ) = p X (x i )p Y (y j ), and so we can estimate p XY (x i, y j ) by ˆp ij = ˆp i ˆp j = n i n j n. Now that we have a set of expected frequencies to compare against our k l observed frequencies, a χ test can be performed. We are using both the row and column sums to estimate our probabilities, and there are k and l of these respectively. So we compare our calculated x statistic against a χ distribution with kl {(k 1) + (l 1)} 1 = (k 1)(l 1) degrees of freedom. Hence the rejection region for a hypothesis test of independence in a k l contingency table at the at the 100α% level is given by R = { } x x > χ (k 1)(l 1),1 α. 14
15 Example An article in International Journal of Sports Psychology (JulySept 1990) evaluated the relationship between physical fitness and stress. 549 people were classified as good, average, or poor fitness, and were also tested for signs of stress (yes or no). The data are shown in the table below. Poor Fitness Average Fitness Good Fitness Stress No stress Is there any relationship between stress and fitness? Under independence we would estimate the expected values to be Poor Fitness Average Fitness Good Fitness Stress No stress Hence the χ statistic is calculated to be X = i (O i E i ) ( ) ( ) = E i = This should be compared with a χ distribution with ( 1) (3 1) = degrees of freedom. χ,0.95 = 5.99, so we have no significant evidence to suggest there is any relationship between fitness and stress. 15
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