Compositional statistical unit testing

How do you unit-test a sampler? Run it a bunch and see whether the histogram looks good—but there’s always some chance that your test will fail randomly even though the sampler is good, or pass randomly despite a bug. And if you try to have more than one test, the chance of random errors goes up. How big of a chance? How much worse does it get? What to do?

I’ve written about this problem before. That was a starting point. In this post, I want to

  • Work out a general formalism for reasoning about the error rates of tests of stochastic software;

  • Work out the rules of composition for tests in this formalism; and

  • Build up to a usable test, namely discrete equality in distribution (one variant where the expected distribution is known and one where it is itself avaiable only as a sampler).

This post is not so good for the math-averse—I need the crutch of precise reasoning to make sure I don’t miss anything important.


Calibrated tests

Suppose we are trying to statistically assess the behavior of some (presumably stochastic) software \(s\). Let \(\S\) be the space of possible true behaviors of \(s\), the software under test. Then a single calibrated test \(T\) is a 5-tuple: \((\tau, S^g, S^b, \alpha, \beta)\) where

  • \(\tau: \S \to \bool\) is a stochastic test procedure returning \(\text{True}\) or \(\text{False}\) (which presumably runs the software under test repeatedly and measures its behavior somehow),
  • \(S^g \subset \S\) is the set of behaviors of \(s\) that are “good”,
  • \(S^b \subset \S\) is the set of behaviors of \(s\) that are “bad”, and
  • \(\alpha \geq 0\) and \(\beta \geq 0\) are error tolerances, such that
  • the false-fail rate is low: if \(s \in S^g\), then \(p(\tau(s) = \text{False}) \leq \alpha\), and
  • the false-pass rate is low: if \(s \in S^b\), then \(p(\tau(s) = \text{True}) \leq \beta\).

For example, consider the situation from my previous post on the subject: \(s\) performs some unknown computation and “fails” with unknown probability \(\pf\). In this case, the number \(\pf\) completely characterizes the behavior of \(s\) that we are interested in, so \(\S = [0,1]\). We wish to bound \(\pf\), so we set \(S^g = [0, 0.1]\). Suppose our test procedure \(\tau\) is “Execute \(s\) independently \(n = 10\) times, and pass if \(s\) ‘fails’ at most 3 of them, inclusive.” Then it’s easy enough to compute that \(\alpha \geq 0.0128\) will satisfy the false-fail rate equation. There are many choices of \(S^b\) and \(\beta\) that will satisfy the false-pass rate equation for this choice of \(\tau\), one of which is \(S^b = [0.7, 1], \beta = 0.011\). As one might imagine, increasing \(n\) can let us decrease \(\alpha\) and \(\beta\), and enlarge \(S^b\).

Note that the above test only makes guarantees about the outcome of \(\tau\) if \(\pf < 0.1\) or \(\pf > 0.7\), but says nothing for the intervening interval. This is a general phenomenon, because \(p(\tau(s) = True)\) is continuous in the tested behavior \(s\). Ergo, if \(\alpha + \beta < 1\), the sets \(S^g\) and \(S^b\) must be separated. If \(S^g\) and \(S^b\) are connected to each other in \(\S\), this implies incomplete coverage: \(S^g \cup S^b \subsetneq \S\).

There are two symmetric ways to think about this gap. One is to say that we, in testing \(s\), know what behavior \(S^g\) is acceptable. In this case, everything else is a “bug” which we would ideally like to distinguish from \(S^g\), but, unless \(S^g\) is both closed and open in \(\S\), we can’t. So we settle for defining \(S^b\) to be the bugs that are “severe enough” that we commit to finding them. In so doing, we balance coverage (a large \(S^b\)) against confidence (small \(\alpha\) and \(\beta\)), and against the computational cost of \(\tau\).

Symmetrically, we could say that we know what behavior \(S^b\) is unacceptable, and define \(S^g\) as the behaviors that are “obviously good” enough to commit to pronouncing them correct. On this view we are balancing the test’s robustness (large \(S^g\)) against confidence and computation.

Calibrated test suites

So much for an isolated calibrated test. How can we compose these things? Like traditional (deterministic) unit tests, we can negate test assertions, and connect multiple assertions with “and” (or “or”). We just need to exercise some care with the resulting good/bad sets and error rates. Unlike tests of deterministic software, it is also meaningful to repeat a calibrated test, as a way of using computation to gain more confidence. Let’s work these out in reverse order.

Amplifying confidence

We can generically trade computation for confidence by repeating a test independently more than once. Indeed, if \(T = (\tau, S^g, S^b, \alpha, \beta)\) is any calibrated test, we can drive down the false-pass rate by forming \(T_\text{and}^k = (\tau^k_\text{and}, S^g, S^b, k\alpha, \beta^k)\). Here, \(\tau^k_\text{and}\) is “Run \(\tau\) independently \(k\) times and return the logical ‘and’ of all the results.” The only way for \(\tau^k_\text{and}\) to pass is if all \(k\) runs of \(\tau\) pass. If this is to be a false pass, i.e., if \(s\) is actually in \(S^b\), the probability of each pass of \(\tau\) is bounded by \(\beta\), which bounds the false-pass rate of \(\tau^k_\text{and}\) by \(\beta^k\), as desired. However, the chance of a false-fail goes up: for \(\tau^k_\text{and}\) to falsely fail, it suffices for at least one of the runs of \(\tau\) to fail even though \(s \in S^g\). The probability of this happening is bounded by \[ \text{if } s \in S^g\quad p(\tau^k_\text{and}(s) = \text{False}) \leq 1 - (1 - \alpha)^k \leq k\alpha. \] The former inequality is tight if \(\alpha\) was a tight bound on the false fail rate of \(\tau\), but the latter is generally conservative. However, it’s simple, and if \(k\alpha\) is, as is usually desired, much less than \(1\), then the quadratic and higher terms in \((1 - \alpha)^k\) are likely to be negligible.

Symmetrically, we can drive down the false-fail rate by forming \(T_\text{or}^k = (\tau^k_\text{or}, S^g, S^b, \alpha^k, k\beta)\), where \(\tau^k_\text{or}\) is “Run \(\tau\) independently \(k\) times and return the logical ‘or’ of all the results.” Since the error rate we are driving down falls exponentially while the other only climbs linearly, we can amplify any non-trivial initial calibrated test \(T\) into one with arbitrarily small \(\alpha\) and \(\beta\). (In fact, with a suitable “\(k\) of \(m\) passes” scheme, any initial error rates satisfying \(\alpha + \beta < 1\) are enough.)

Composing tests

The calibrated test formalism gets more interesting when we consider combining different tests. Suppose we have \(T_1 = (\tau_1, S^g_1, S^b_1, \alpha_1, \beta_1)\) and \(T_2 = (\tau_2, S^g_2, S^b_2, \alpha_2, \beta_2)\). Then we can get a more specific test that’s still calibrated. Let \(\tau_1 \cdot \tau_2\) be the testing procedure that runs \(\tau_1\) and \(\tau_2\) independently and returns the logical “and” of the answers. Then \[ T_1 \cdot T_2 = (\tau_1 \cdot \tau_2, S^g_1 \cap S^g_2, S^b_1 \cup S^b_2, \alpha_1 + \alpha_2, \max(\beta_1, \beta_2)) \] is a calibrated test. Why? If the software is good according to both \(T_1\) and \(T_2\), the false fail rate is bounded by \[ p((\tau_1 \cdot \tau_2)(s) = \text{False}) \leq 1 - (1 - \alpha_1)(1 - \alpha_2) \leq \alpha_1 + \alpha_2.\]

Conversely, if the software is bad according to either \(T_1\) or \(T_2\), then it has to obey one or the other false-pass rate equation, so \(\max(\beta_1, \beta_2)\) bounds the overall false-pass rate. Note, by the way, that if we stick to the tight bound \(1 - (1 - \alpha_1)(1 - \alpha_2)\) rather than the simple \(\alpha_1 + \alpha_2\), the \(\cdot\) operation on tests becomes associative.

What have we gained? By being a little careful with the confidences, we can form compound tests that ensure all of a set of desirable properties \(S^g_i\) obtain together, or, equivalently, weed out any of a set of severe bugs \(S^b_i\). In other words, we can build up complex specifications of “correctness” by intersecting simpler specifications \(S^g_i\), while maintaining reasonable coverage through unioning the corresponding unacceptable sets \(S^b_i\). This is the basis on which developers of stochastic software \(s\) can construct a test suite—if they take care to tighten the \(\alpha\) bounds of the tests from time to time so that the \(\alpha\) of the overall test suite stays within acceptable limits.

For the record, the symmetric definition of \(\tau_1 + \tau_2\) as the logical “or” of \(\tau_1\) and \(\tau_2\) gives us \[ T_1 + T_2 = (\tau_1 + \tau_2, S^g_1 \cup S^g_2, S^b_1 \cap S^b_2, \max(\alpha_1, \alpha_2), \beta_1 + \beta_2), \] which is a way of building up compound unacceptable behaviors by intersecting simpler ones.

Negating tests

Finally, no notion of a test suite is complete without the idea of a test that is known to fail, because of a known bug in the underlying software that just hasn’t been fixed yet. The corresponding concept here is test negation: given a calibrated test \(T = (\tau, S^g, S^b, \alpha, \beta)\), we can invert the sense of \(\tau\) by running it once and negating the answer. The resulting \(\tilde \tau\) gives the calibrated test \[ \tilde T = (\tilde \tau, S^b, S^g, \beta, \alpha),\] which makes the same distinction as \(T\) with the same confidence, but with the reverse sense.

This does not fully cover the role “expected failure” plays in deterministic testing. The latter has the property that if some test fails for an unknown reason that one does not currently have time to deal with, one can mark that test “expected failure”, get the test suite to pass that way, and then return to investigating the problem later. Negating a statistical test does not reliably obtain the same effect, because the software \(s\) could be in the dead zone \(\S - (S^g \cup S^b)\), where neither \(T\) nor \(\tilde T\) will pass reliably. Sadly, there is nothing that can be done at the level of opaque calibrated tests in this situation. Perhaps a practical testing framework should have an additional marking whose effect is “run this test (to check that it completes without crashing) but ignore the answer” to accommodate this situation.

Usable tests

Let us now turn our attention to using this machinery to build the simplest actually usable calibrated tests: discrete equality in distribution. We will get two variants: the general version, where the “expected” distribution is itself given by a sampler, and a more efficient version where the “expected” distribution is an explicit list of objects and their probabilities. The latter is also called a “goodness of fit” test.

The student of statistics will recognize that both of these tasks have well-known half-calibrated tests (the \(\chi^2\) family is popular). I call the standard tests “half”-calibrated because, while their false-fail rates are well-understood, I am not aware of any characterization of their false-pass rates. The obstacle, presumably, is that such a characterization requires understanding the behavior of the test statistic for any possible “bad” state of the software under test.

A Building Block

Let us start with a two-sample inequality in probability test, as this will let us build both of our promised compound tests. To wit, suppose our software \(s\) under test provides two operations, \(s_1\) and \(s_2\), each of which returns true or false. We wish to check that \(p_1 = p(s_1 = \text{True}) \leq p_2 = p(s_2 = \text{True})\).

Formally, our software \(s\) is characterized completely by the two numbers \(p_1\) and \(p_2\), so \(\S = [0,1] \times [0,1]\). Our desired “good” region is \(S^g = \{p_1 \leq p_2\}\). \(S^b\) must be disconnected from \(S^g\), but it would be nice to have a parameter by which we can let it approach arbitrarily close; so let’s pick \(S^b = \{ p_1 \geq p_2 + \eps \}\) for some \(\eps > 0\).

What can be do for a testing procedure \(\tau\)? Not much, really. The simplest option1 is to run \(s_1\) some number \(n_1\) times, run \(s_2\) \(n_2\) times, and count the numbers \(k_1\) and \(k_2\) that they respectively produce True. Having obtained those results, use some decision rule to emit True or False from the test as a function of \(k_1\) and \(k_2\), and then calibrate \(\alpha\) and \(\beta\) by computing or bounding the probability of False if \(p_1 \leq p_2\), and of True if \(p_1 \geq p_2 + \eps\).

I don’t have any special insight on the best way to choose such a decision rule, but here’s one that will do: \[ \tau(n_1, n_2, \eps)(s) = \begin{cases} \text{True} & \text{if } \frac{k_1}{n_1} \leq \frac{k_2}{n_2} + \frac{\eps}{2} \\ \text{False} & \text{otherwise}. \end{cases} \]

What false-fail rate does this have? Well, if \(s\) is actually characterized by \(p_1, p_2\), then the probability of any given outcome \(k_1, k_2\) is \[ p(k_1, k_2|p_1, p_2) = {n_1 \choose k_1} p_1^{k_1} (1 - p_1)^{n_1 - k_1} {n_2 \choose k_2} p_2^{k_2} (1 - p_2)^{n_2 - k_2}. \] The probability of failure is the probability that any of the outcomes leading to False will occur, and the worst-case false-fail rate is that probability for the \(p_1 \leq p_2\) that maximize it: \[ \begin{align*} \alpha & = \max_{p_1 \leq p_2} \left[ \sum_{\frac{k_1}{n_1} > \frac{k_2}{n_2} + \frac{\eps}{2}} {n_1 \choose k_1} p_1^{k_1} (1 - p_1)^{n_1 - k_1} {n_2 \choose k_2} p_2^{k_2} (1 - p_2)^{n_2 - k_2} \right] \\ & = \max_{p_1 \leq p_2} \left[ \sum_{k_2} {n_2 \choose k_2} p_2^{k_2} (1 - p_2)^{n_2 - k_2} \left( \sum_{\frac{k_1}{n_1} > \frac{k_2}{n_2} + \frac{\eps}{2}} {n_1 \choose k_1} p_1^{k_1} (1 - p_1)^{n_1 - k_1} \right) \right]. \\ \end{align*} \]

The inner sum is convenient, being the survivor function of the binomial distribution of \(n_1\) trials with success rate \(p_1\). Since the binomial survivor function is at every point monotonic in the success probability, it’s safe to say that for any given \(p_2\), the maximum over \(p_1\) is at the highest value admissible by the constraint, in this case \(p_1 = p_2\). This reduces our optimization problem to one dimension.

The false-pass rate is similar: \[ \beta = \max_{p_1 \geq p_2 + \eps} \left[ \sum_{k_2} {n_2 \choose k_2} p_2^{k_2} (1 - p_2)^{n_2 - k_2} \left( \sum_{\frac{k_1}{n_1} \leq \frac{k_2}{n_2} + \frac{\eps}{2}} {n_1 \choose k_1} p_1^{k_1} (1 - p_1)^{n_1 - k_1} \right) \right], \] with a similar simplification due to the inner sum being a binomial cumulative distribution function, which is maximized by minimizing \(p_1\).

The above optimizations are not very hard to solve. Here are a few computed error rates for tests with given \(\eps\), \(n_1\), and \(n_2\) parameters:

\(\eps\) \(n_1\) \(n_2\) \(\alpha\) \(\beta\)
0.20 30 30 0.183155 0.253557
0.20 100 100 0.072367 0.082141
0.20 200 200 0.020901 0.023023
0.20 271 271 0.009036 0.009910
0.10 30 30 0.349442 0.347353
0.10 100 100 0.218377 0.260811
0.10 300 300 0.102815 0.117143
0.10 1000 1000 0.011947 0.013132
0.10 1085 1085 0.009634 0.009974
0.05 30 30 0.448711 0.397353
0.05 100 100 0.361886 0.361332
0.05 300 300 0.270163 0.269769
0.05 1000 1000 0.127058 0.136325
0.05 3000 3000 0.025816 0.026889
0.05 4334 4334 0.009880 0.009995

Two-sample equality in distribution

With that capability in hand, we can now ask for something practically usable that I don’t know any other way to get: a calibrated test that two discrete samplers are sampling from the same distribution. To wit, suppose \(s\) now consists of two programs \(s_1\) and \(s_2\), each of which randomly emits one of \(D\) tokens. We wish to test that \(s_1\) and \(s_2\) have the same probability of emitting each token.

This can be accomplished with a conjunction of \(D\) of the two-sample inequality tests described above: just assert that the probability of \(s_1\) generating each token \(d\) is no more than the probability of \(s_2\) generating the same token: \[ \text{And}_{d \in D} \big[p(s_1 = d) \leq p(s_2 = d)\big]. \] The reverse inequalities follow by conservation of belief—the total probability in both distributions must be 1.

One could use this to, for example, design a calibrated unit test for, say, a Chinese Restaurant Process sampler. The CRP is a probability distribution on partitions. There are 18 possible partitions of a set of 4 distinct objects into clusters. So, one could design a unit test that checks whether two purported samplers for the CRP distribution on 4 objects in fact agree. In this style, that test would check, independently for each of the 18 possible partitions, that the probability of drawing that one from program 1 is no more than the probability of drawing it from program 2. If one designed each test to make \(n\) trials, one would need \(18n\) trials in total, and, using the above single-inequality procedure, one could obtain the following computation-precision-confidence trade off:

\(\eps\) \(n\) \(18n\) \(\alpha\) \(\beta\)
0.20 30 540 0.973787 0.253557
0.20 100 1800 0.741313 0.082141
0.20 300 5400 0.109433 0.007154
0.20 530 9540 0.009010 0.000531
0.10 30 540 0.999564 0.347353
0.10 100 1800 0.988144 0.260811
0.10 300 5400 0.858134 0.117143
0.10 1000 18000 0.194546 0.013132
0.10 2120 38160 0.009576 0.000572
0.05 30 540 0.999978 0.397353
0.05 100 1800 0.999692 0.361332
0.05 300 5400 0.996548 0.269769
0.05 1000 18000 0.913356 0.136325
0.05 3000 54000 0.375490 0.026889
0.05 8480 152640 0.009889 0.000573

If those trial runs look a bit large, there are couple reasons:

  • Each individual test of the 18 is pessimistic, guaranteeing its error rates even if the probability of that particular partition is the most confusing possible, namely close to 1/2. But, of course, they can’t all be close to 1/2. A directly synthetic test like the \(\chi^2\) test of independence is likely to be much more efficient. Learning how to calibrate its false pass rate would be a useful advance.

  • Relatedly, the design outlined here wastes computation on assuring the independence of the individual tests. Namely, when one is measuring the frequency of some partition, one could presumably use those samples to measure the frequency of each of the 17 others as well. I think it would be fruitful to think through what restrictions, if any, there are on sharing samples across tests of different facets of the same sampler.

  • In this particular design, I arbitrarily chose that \(\tau(n_1, n_2, \eps) = \text{True}\) when \(k_1/n_1 \leq k_2/n_2 + \eps/2\). That threshold balances the false fail and false pass rates for each individual test, but the compound test, being an “and”, has a substantially higher false fail rate than each component, but the same false pass rate. Therefore, some more efficiency can be obtained by making the threshold larger than \(k_2/n_2 + \eps/2\). That way, the computation used to ensure the false fail rate is acceptable is not wasted on driving the false pass rate lower than needed.

One-sample Equality in Distribution

We can actually gain a lot of computational efficiency if we know the expected distribution analytically. Then we can use a one-sample test on each dimension, and save many trials by not having to account for too many hard-to-assess probabilities that are near 0.5.

In the case of the Chinese Restaurant Process, the analytic distribution on partitions is known. For example, if the concentration parameter is 0.5, the 18 partitions have these probabilities:

Partition Probability
[1, 1, 1, 1] 0.457142857143
[1, 1, 1, 2] 0.0761904761905
[1, 1, 2, 1] 0.0761904761905
[1, 1, 2, 2] 0.0380952380952
[1, 1, 2, 3] 0.0190476190476
[1, 2, 1, 1] 0.0761904761905
[1, 2, 1, 2] 0.0380952380952
[1, 2, 1, 3] 0.0190476190476
[1, 2, 2, 1] 0.0380952380952
[1, 2, 2, 2] 0.0761904761905
[1, 2, 2, 3] 0.0190476190476
[1, 2, 3, 1] 0.0190476190476
[1, 2, 3, 2] 0.0190476190476
[1, 2, 3, 3] 0.0190476190476
[1, 2, 3, 4] 0.00952380952381

Using the one-sample test design function from my previous post, we can design 18 tests corresponding to these probabilities. These designs will naturally spend less computation on the low-probability partitions, because those are easier to reliably falsify. With that design, we can get to \(0.01\) error rates with these total trial counts:

\(\eps\) Trials
0.2 1244
0.1 3894
0.05 12813

A dramatic improvement over the two-sample situation.


I think the composable structure of calibrated tests forms a sufficient backbone for an intellectually sound unit testing framework for stochastic software. Of course, a few more things do need to be worked out before one can become practical. Some come to mind offhand:

  1. It would be nice to get more efficient calibrated tests (perhaps in the \(\chi^2\) style) for equality in distribution than those presented here. I just started with these to prove that it could be done, and to demonstrate how the compositions tend to work out.

  2. Continuous distributions can be attacked by binning, but it would be more satisfying to calibrate something like the Kolmogorov-Smirnov test.

  3. Any library would do well to provide tests of nearness (as opposed to equality) in distribution, with the distance bound given either as an explicit number or implicitly as the distance between two other distributions. This should be particularly useful for testing approximate inference algorithms, which promise only to move a distribution toward a goal, not to actually get there. Designing such tests directly could be substantially more efficient than composing them out of 1-D probability inequalities.

  4. There is an interesting design space of continuous integration tools, since now there is confidence to be gained by rerunning a test that already passed, or increasing its computation budget internally.

  5. I’ve presented forward computations, deriving the best obtainable error bounds from a given test procedure. I think a tool could become substantially more practical if it automated the experiment design problem: The user tells the framework the \(S^g, S^b, \alpha\), and \(\beta\) they want, and the framework finds parameters for the testing procedure(s) \(\tau\) to meet those requirements while using as little computation as it can. This becomes very interesting when compound tests are involved, because the framework could (within limits) squeeze more confidence out of cheaper individual tests, and go a bit easier on expensive ones, and still get its overall error rate bounds.

But, that’s all!


  1. Not the only option, as one can imagine choosing \(n_1\) and \(n_2\) dynamically, as results from previous trials come in. Such a technique may be a way to save computation, but a sound analysis is beyond the scope of this blog post.