Introduction

I have been recently reviewing and rewriting a paper for publication that I first wrote back in 2011. The paper (Wallis forthcoming) concerns the problem of how we test whether repeated runs of the same experiment obtain essentially the same results, i.e. results are not significantly different from each other.

These meta-tests can be used to test an experiment for replication: if you repeat an experiment and obtain significantly different results on the first repetition, then, with a 1% error level, you can say there is a 99% chance that the experiment is not replicable.

These tests have other applications. You might be wishing to compare your results with those of others in the literature, compare results with different operationalisation (definitions of variables), or just compare results obtained with different data – such as comparing a grammatical distribution observed in speech with that found within writing.

The design of tests for this purpose is addressed within the t-testing ANOVA community, where tests are applied to continuously-valued variables. The solution concerns a particular version of an ANOVA, called “the test for interaction in a factorial analysis of variance” (Sheskin 1997: 489).

However, anyone using data expressed as discrete alternatives (A, B, C etc) has a problem: the classical literature does not explain what you should do.

Gradient and point tests

gradient-point-test

Figure 1: Point tests (A) and gradient tests (B), from Wallis (forthcoming).

The rewrite of the paper caused me to distinguish between two types of tests: ‘point tests’, which I describe below, and ‘gradient tests’.

These tests can be used to compare results drawn from 2 × 2 or r × c χ² tests for homogeneity (also known as tests for independence). This is the most common type of contingency test, which can be computed using Fisher’s exact method or as a Newcombe-Wilson difference interval.

  • A gradient test (B) evaluates whether the gradient or difference between point 1 and point 2 differs between runs of an experiment, dp₁ – p₂. This concerns whether claims about the rate of change, or size of effect, observed are replicable. Gradient tests can be extended, with increasing degrees of freedom, into tests comparing patterns of effect.
  • A point test (A) simply asks whether data at either point, evaluated separately, differs between experimental runs. This concerns whether single observations, such as p₁, are replicable. Point tests can be extended into ‘multi-point’ tests, which we discuss below.

Point tests only apply to homogeneity data. If you wish to compare outcomes from goodness of fit tests, you need a version of the gradient test, to compare differences from an expected Pdp₁ – P. Since different data sets may have different expected P, a distinct ‘point test for goodness of fit’ would be meaningless.

The earlier version of the paper, which has been published on this blog since its launch 2012, focused on gradient tests. The possibility of carrying out a point test was mentioned in passing. In this blog post I want to focus on point tests.

The obvious problem with gradient tests is that two experimental runs might obtain the same gradient but in fact be very different in start and end points. Consider the following graph.

equal-gradient

Figure 2: Why we need two different types of test: (almost) equal gradients but unequal points.

Point tests

The data in Figure 1 is calculated from two 2 × 2 tables drawn from a paper by Aarts, Close and Wallis (2013).

Note: To obtain Figure 2, I simply replaced one frequency in the first table: 46 with 100. The data is also found on the 2×2 homogeneity tab in this Excel spreadsheet, which contains a wide range of separability tests.

To make our exposition clearer, Table 1 uses the same format as in the Excel spreadsheet (with the dependent variable distributed vertically) rather than the format in the paper.

spoken LLC
(1960s)
ICE-GB
(1990s)
Total
shall 124 46 170
will 501 544 1,045
Total 625 590 1,215
written LOB
(1960s)
FLOB
(1990s)
Total
shall 355 200 555
will 2,798 2,723 5,521
Total 3,153 2,923 6,076

Frequency data for the choice modal shall out of the choice shall vs. will, various sources, from Aarts et al. (2013).

Aarts et al. carried out 2 × 2 homogeneity tests for the two tables separately. These test whether modal shall declines as a proportion of the modal shall/will alternation between the two time points. In other words, we compare LLC with ICE-GB data, and LOB with FLOB data.

To carry out a point test we simply rotate the test 90 degrees, e.g. to compare data at the 1960s point we compare LLC with LOB.

As I have explained elsewhere (Wallis 2013), there are a number of different methods for carrying out this comparison.

These include:

  1. The z test for two independent proportions (Sheskin 1997: 226).
  2. The Newcombe-Wilson interval test (Newcombe 1998).
  3. The 2 × 2 χ² test for homogeneity (independence).

These are all standard tests and each is discussed in papers and elsewhere on this blog.

The advantage of the third approach is that it is extensible to c-way multinomial observations by using a 2 × c χ² test.

The multi-point test

The tests listed above can be used to compare the 1960s and 1990s intervals in Figure 1 separately.

However, in many cases it would be helpful to have a method that evaluated both pairs of observations in a single test. This can be generalised to a series of r observations. To do this, in (Wallis forthcoming) I propose what I call a multi-point test.

We generalise the χ² formula by summing over i = 1..r:

  • χd² = ∑χ²(i)

where χ²(i) represents the χ² score for homogeneity for each set of data at position i in the distribution.

This test has r × df(i) degrees of freedom, where df(i) is the degrees of freedom for each χ² point test. So, in the worked example we have seen, the summed test has two degrees of freedom:

spoken LLC
(1960s)
ICE-GB
(1990s)
Total
shall 124 46 170
will 501 544 1,045
Total 625 590 1,215
written LOB
(1960s)
FLOB
(1990s)
Total
shall 355 200 555
will 2,798 2,723 5,521
Total 3,153 2,923 6,076
χ² 34.6906 0.6865 35.3772

Applying the generalised point test calculation to the table above. χ² = 35.38 is significant with 2 degrees of freedom and α = 0.05.

Since the computation sums independently-calculated χ² scores, each score may be individually considered for significant difference (with df(i) degrees of freedom). Hence we can see above the large score for the 1960s data (individually significant) and the small score for 1990s (individually non-significant).

Note: Whereas χ² is generally associative (non-directional), the summed equation (χd²) is not. Nor is this computation the same as a 3 dimensional test (t × r × c). Variables are treated differently.

  • The multi-point test factors out variation between tests over the independent variable (in this instance: time). This means that if there is a lot more data in one table at a particular time period, this fact does not skew the results.
  • On the other hand, it does not factor out variation over the dependent variable – after all, this is precisely what we wish to examine!

Naturally, like the point test, this test may be generalised to multinomial observations.

A Newcombe-Wilson multi-point test

An alternative multi-point test for binomial (two-way) variables employs a sum of χ² values abstracted from Newcombe-Wilson tests.

  1. Carry out Newcombe-Wilson tests for each point test i at a given error level α, obtaining Di, Wi⁻ and Wi⁺.
  2. Identify the inner interval width Wi for each test:
    • if D< 0, Wi = Wi⁻; WiWi⁺ otherwise.
  3. Use the difference Di and inner interval Wi to compute χ² scores:
    • χ²(i) = (Di . zα/2 / Wi)².

It is then possible to sum χ²(i) as before.

Using the data in the worked example we obtain:

1960s: Di = 0.0858, Wi⁻ = -0.0347 and Wi⁺ = 0.0316 (significant).
1990s: Di = 0.0095, Wi⁻ = -0.0194 and Wi⁺ = 0.0159 (ns).

Since Di is positive in both cases, we use the upper interval width each time. This gives us χ² scores of 28.4076 and 1.3769 respectively, which obtains a sum of 29.78. Compared to the first method above, this approach tends to downplay extreme differences.

In conclusion

The point test and the additive generalisation of this test into a ‘multi-point test’ represent a method of contrasting multiple runs of the same experiment, comparing observed changes in different subcorpora or genres, or examine the empirical effect of changing definitions of variables.

These tests consider the null hypothesis that individual observations are not different; or, in the multi-point case, that in general the observations are not different.

  • They do not evaluate the gradient between points or the size of effect. If we wish to compare sizes of effect we would need to use one of the methods for this purpose described in (Wallis forthcoming).
  • The method only applies to comparing tests for homogeneity (independence). To compare goodness of fit data, a different approach is required (also described in Wallis forthcoming).

Nonetheless, these tests are useful meta-tests that build on classical Pearson χ² tests, and they are useful tools in our analytical armoury.

See also

References

Sheskin, D.J. 1997. Handbook of Parametric and Nonparametric Statistical Procedures. Boca Raton, Fl: CRC Press.

Newcombe, R.G. 1998. Interval estimation for the difference between independent proportions: comparison of eleven methods. Statistics in Medicine 17: 873-890.

Wallis, S.A. 2013. z-squared: the origin and application of χ². Journal of Quantitative Linguistics 20:4, 350-378. » Post

Wallis, S.A. forthcoming (first published 2011). Comparing χ² tables for separability of distribution and effect. London: Survey of English Usage. » Post

Detecting direction in interaction evidence

IntroductionPaper (PDF)

I have previously argued (Wallis 2014) that interaction evidence is the most fruitful type of corpus linguistics evidence for grammatical research (and doubtless for many other areas of linguistics).

Frequency evidence, which we can write as p(x), the probability of x occurring, concerns itself simply with the overall distribution of a linguistic phenomenon x – such as whether informal written English has a higher proportion of interrogative clauses than formal written English. In order to calculate frequency evidence we must define x, i.e. decide how to identify interrogative clauses. We must also pick an appropriate baseline n for this evaluation, i.e. we need to decide whether to use words, clauses, or any other structure to identify locations where an interrogative clause may occur.

Interaction evidence is different. It is a statistical correlation between a decision that a writer or speaker makes at one part of a text, which we will label point A, and a decision at another part, point B. The idea is shown schematically in Figure 1. A and B are separate ‘decision points’ in a given relationship (e.g. lexical adjacency), which can be also considered as ‘variables’.

Figure 1: Associative inference from lexico-grammatical choice variable A to variable B (sketch).

Figure 1: Associative inference from lexico-grammatical choice variable A to variable B (sketch).

This class of evidence is used in a wide range of computational algorithms. These include collocation methods, part-of-speech taggers, and probabilistic parsers. Despite the promise of interaction evidence, the majority of corpus studies tend to consist of discussions of frequency differences and distributions.

In this paper I want to look at applications of interaction evidence which are made more-or-less at the same time by the same speaker/writer. In such circumstances we cannot be sure that just because B follows A in the text, the decision relating to B was made after the decision at A. Continue reading

UCL Summer School in English Corpus Linguistics 2017

I am pleased to announce the fifth annual Summer School in English Corpus Linguistics to be held at University College London from 5-7 July.

The Summer School is a short three-day intensive course aimed at PhD-level students and researchers who wish to get to grips with Corpus Linguistics. Numbers are deliberately limited on a first-come, first-served basis. You will be taught in a small group by a teaching team.

Each day begins with a theory lecture, followed by a guided hands-on workshop with corpora, and a more self-directed and supported practical session in the afternoon.

W9A8081
Continue reading

The replication crisis: what does it mean for corpus linguistics?

Introduction

Over the last year, the field of psychology has been rocked by a major public dispute about statistics. This concerns the failure of claims in papers, published in top psychological journals, to replicate.

Replication is a big deal: if you publish a correlation between variable X and variable Y – that there is an increase in the use of the progressive over time, say, and that increase is statistically significant, you expect that this finding would be replicated were the experiment repeated.

I would strongly recommend Andrew Gelman’s brief history of the developing crisis in psychology. It is not necessary to agree with everything he says (personally, I find little to disagree with, although his argument is challenging) to recognise that he describes a serious problem here.

There may be more than one reason why published studies have failed to obtain compatible results on repetition, and so it is worth sifting these out.

In this blog post, what I want to do is try to explore what this replication crisis is – is it one problem, or several? – and then turn to what solutions might be available and what the implications are for corpus linguistics. Continue reading

The variance of Binomial distributions

Introduction

Recently I’ve been working on a problem that besets researchers in corpus linguistics who work with samples which are not drawn randomly from the population but rather are taken from a series of sub-samples. These sub-samples (in our case, texts) may be randomly drawn, but we cannot say the same for any two cases drawn from the same sub-sample. It stands to reason that two cases taken from the same sub-sample are more likely to share a characteristic under study than two cases drawn entirely at random. I introduce the paper elsewhere on my blog.

In this post I want to focus on an interesting and non-trivial result I needed to address along the way. This concerns the concept of variance as it applies to a Binomial distribution.

Most students are familiar with the concept of variance as it applies to a Gaussian (Normal) distribution. A Normal distribution is a continuous symmetric ‘bell-curve’ distribution defined by two variables, the mean and the standard deviation (the square root of the variance). The mean specifies the position of the centre of the distribution and the standard deviation specifies the width of the distribution.

Common statistical methods on Binomial variables, from χ² tests to line fitting, employ a further step. They approximate the Binomial distribution to the Normal distribution. They say, although we know this variable is Binomially distributed, let us assume the distribution is approximately Normal. The variance of the Binomial distribution becomes the variance of the equivalent Normal distribution.

In this methodological tradition, the variance of the Binomial distribution loses its meaning with respect to the Binomial distribution itself. It seems to be only valuable insofar as it allows us to parameterise the equivalent Normal distribution.

What I want to argue is that in fact, the concept of the variance of a Binomial distribution is important in its own right, and we need to understand it with respect to the Binomial distribution, not the Normal distribution. Sometimes it is not necessary to approximate the Binomial to the Normal, and if we can avoid this approximation our results are likely to be stronger as a result.

Continue reading

Adapting variance for random-text sampling

Introduction Paper (PDF)

Conventional stochastic methods based on the Binomial distribution rely on a standard model of random sampling whereby freely-varying instances of a phenomenon under study can be said to be drawn randomly and independently from an infinite population of instances.

These methods include confidence intervals and contingency tests (including multinomial tests), whether computed by Fisher’s exact method or variants of log-likelihood, χ², or the Wilson score interval (Wallis 2013). These methods are also at the core of others. The Normal approximation to the Binomial allows us to compute a notion of the variance of the distribution, and is to be found in line fitting and other generalisations.

In many empirical disciplines, samples are rarely drawn “randomly” from the population in a literal sense. Medical research tends to sample available volunteers rather than names compulsorily called up from electoral or medical records. However, provided that researchers are aware that their random sample is limited by the sampling method, and draw conclusions accordingly, such limitations are generally considered acceptable. Obtaining consent is occasionally a problematic experimental bias; actually recruiting relevant individuals is a more common problem.

However, in a number of disciplines, including corpus linguistics, samples are not drawn randomly from a population of independent instances, but instead consist of randomly-obtained contiguous subsamples. In corpus linguistics, these subsamples are drawn from coherent passages or transcribed recordings, generically termed ‘texts’. In this sampling regime, whereas any pair of instances in independent subsamples satisfy the independent-sampling requirement, pairs of instances in the same subsample are likely to be co-dependent to some degree.

To take a corpus linguistics example, a pair of grammatical clauses in the same text passage are more likely to share characteristics than a pair of clauses in two entirely independent passages. Similarly, epidemiological research often involves “cluster-based sampling”, whereby each subsample cluster is drawn from a particular location, family nexus, etc. Again, it is more likely that neighbours or family members share a characteristic under study than random individuals.

If the random-sampling assumption is undermined, a number of questions arise.

  • Are statistical methods employing this random-sample assumption simply invalid on data of this type, or do they gracefully degrade?
  • Do we have to employ very different tests, as some researchers have suggested, or can existing tests be modified in some way?
  • Can we measure the degree to which instances drawn from the same subsample are interdependent? This would help us determine both the scale of the problem and arrive at a potential solution to take this interdependence into account.
  • Would revised methods only affect the degree of certainty of an observed score (variance, confidence intervals, etc.), or might they also affect the best estimate of the observation itself (proportions or probability scores)?

Continue reading

Impossible logistic multinomials

Introduction

Recently, a number of linguists have begun to question the wisdom of assuming that linguistic change tends to follow an ‘S-curve’ or more properly, logistic, pattern. For example, Nevalianen (2015) offers a series of empirical observations that show that whereas data sometimes follows a continuous ‘S’, frequently this does not happen. In this short article I try to explain why this result should not be surprising.

The fundamental assumption of logistic regression is that a probability representing a true fraction, or share, of a quantity undergoing a continuous process of change by default follows a logistic pattern. This is a reasonable assumption in certain limited circumstances because an ‘S-curve’ is mathematically analogous to a straight line (cf. Newton’s first law of motion).

Regression is a set of computational methods that attempts to find the closest match between an observed set of data and a function, such as a straight line, a polynomial, a power curve or, in this case, an S-curve. We say that the logistic curve is the underlying model we expect data to be matched against (regressed to). In another post, I comment on the feasibility of employing Wilson score intervals in an efficient logistic regression algorithm.

We have already noted that change is assumed to be continuous, which implies that the input variable (x) is real and linear, such as time (and not e.g. probabilistic). In this post we discuss different outcome variable types. What are the ‘limited circumstances’ in which logistic regression is mathematically coherent?

  • We assume probabilities are free to vary from 0 to 1.
  • The envelope of variation must be constant, i.e. it must always be possible for an observed probability to reach 1.

Taken together this also means that probabilities are Binomial, not multinomial. Let us discuss what this implies. Continue reading