mathstoshare

Category: Probability

  • Poisson approximations to the negative binomial distribution

    This post is an introduction to the negative binomial distribution and a discussion of different ways of approximating the negative binomial distribution.

    The negative binomial distribution describes the number of times a coin lands on tails before a certain number of heads are recorded. The distribution depends on two parameters \(p\) and \(r\). The parameter \(p\) is the probability that the coin lands on heads and \(r\) is the number of heads. If \(X\) has the negative binomial distribution, then \(X = x\) means in the first \(x+r-1\) tosses of the coin, there were \(r-1\) heads and that toss number \(x+r\) was a head. This means that the probability that \(X=x\) is given by

    \(\displaystyle{f(x) = \binom{x+r-1}{r-1}p^{r}\left(1-p\right)^x}\)

    Here is a plot of the function \(f(x)\) for different values of \(r\) and \(p\).

    Poisson approximations

    When the parameter \(r\) is large and \(p\) is close to one, the negative binomial distribution can be approximated by a Poisson distribution. More formally, suppose that \(r(1-p)=\lambda\) for some positive real number \(\lambda\). If \(r\) is large then, the negative binomial random variable with parameters \(p\) and \(r\), converges to a Poisson random variable with parameter \(\lambda\). This is illustrated in the picture below where three negative binomial distributions with \(r(1-p)=5\) approach the Poisson distribution with \(\lambda =5\).

    Total variation distance is a common way to measure the distance between two discrete probability distributions. The log-log plot below shows that the error from the Poisson approximation is on the order of \(1/r\) and that the error is bigger if the limiting value of \(r(1-p)\) is larger.

    It turns out that is is possible to get a more accurate approximation by using a different Poisson distribution. In the first approximation, we used a Poisson random variable with mean \(\lambda = r(1-p)\). However, the mean of the negative binomial distribution is \(r(1-p)/p\). This suggests that we can get a better approximation by setting \(\lambda = r(1-p)/p\).

    The change from \(\lambda = r(1-p)\) to \(\lambda = r(1-p)/p\) is a small because \(p \approx 1\). However, this small change gives a much better approximation, especially for larger values of \(r(1-p)\). The below plot shows that both approximations have errors on the order of \(1/r\), but the constant for the second approximation is much better.

    Second order accurate approximation

    It is possible to further improve the Poisson approximation by using a Gram–Charlier expansion. A Gram–Charlier approximation for the Poisson distribution is given in this paper.1 The approximation is

    \(\displaystyle{f_{GC}(x) = P_\lambda(x) – \frac{1}{2}(1-p)\left((x-\lambda)P_\lambda(x)-(x-1-\lambda)P_\lambda(x-1)\right)},\)

    where \(\lambda = \frac{k(1-p)}{p}\) as in the second Poisson approximation and \(P_\lambda(x)\) is the Poisson pmf evaluated at \(x\).

    The Gram–Charlier expansion is considerably more accurate than either Poisson approximation. The errors are on the order of \(1/r^2\). This higher accuracy means that the error curves for the Gram–Charlier expansion has a steeper slope.

    1. The approximation is given in equation (4) of the paper and is stated in terms of the CDF instead of the PMF. The equation also contains a small typo, it should say \(\frac{1}{2}q\) instead of \(\frac{1}{2}p\). ↩︎

  • “Uniformly random”

    The term “uniformly random” sounds like a contradiction. How can the word “uniform” be used to describe anything that’s random? Uniformly random actually has a precise meaning, and, in a sense, means “as random as possible.” I’ll explain this with an example about shuffling card.

    Shuffling cards

    Suppose I have a deck of ten cards labeled 1 through 10. Initially, the cards are face down and in perfect order. The card labeled 10 is on top of the deck. The card labeled 9 is second from the top, and so on down to the card labeled 1. The cards are definitely not random.

    Next, I generate a random number between 1 and 10. I then find the card with the corresponding label and put it face down on top of the deck. The cards are now somewhat random. The number on top could anything, but the rest of the cards are still in order. The cards are random but they are not uniformly random.

    Now suppose that I keep generating random numbers and moving cards to the top of the deck. Each time I do this, the cards get more random. Eventually (after about 30 moves1) the cards will be really jumbled up. Even if you knew the first few cards, it would be hard to predict the order of the remaining ones. Once the cards are really shuffled, they are uniformly random.

    Uniformly random

    A deck of cards is uniformly random if each of the possible arrangements of the cards are equally likely. After only moving one card, the deck of cards is not uniformly random. This is because there are only 10 possible arrangements of the deck. Once the deck is well-shuffled, all of the 3,628,800 possible arrangements are equally likely.

    In general, something is uniformly random if each possibility is equally likely. So the outcome of rolling a fair 6-sided die is uniformly random, but rolling a loaded die is not. The word “uniform” refers to the chance of each possibility (1/6 for each side of the die). These chances are all the same and “uniform”.

    This is why uniformly random can mean “as random as possible.” If you are using a fair die or a well-shuffled deck, there are no biases in the outcome. Mathematically, this means you can’t predict the outcome.

    Communicating research

    The inspiration for this post came from a conversation I had earlier in the week. I was telling someone about my research. As an example, I talked about how long it takes for a deck of cards to become uniformly random. They quickly stopped me and asked how the two words could ever go together. It was a good point! I use the words uniformly random all the time and had never realized this contradiction.2 It was a good reminder about the challenge of clear communication.

    Footnotes

    1. The number of moves it takes for the deck to well-shuffled is actually random. But on average it takes around 30 moves. For the mathematical details, see Example 1 in Shuffling Cards and Stopping Times by David Aldous and Persi Diaconis. ↩︎
    2. Of the six posts I published last year, five contain the word uniform! ↩︎

  • Understanding the Ratio of Uniforms Distribution

    The ratio of uniforms distribution is a useful distribution for rejection sampling. It gives a simple and fast way to sample from discrete distributions like the hypergeometric distribution1. To use the ratio of uniforms distribution in rejection sampling, we need to know the distributions density. This post summarizes some properties of the ratio of uniforms distribution and computes its density.

    The ratio of uniforms distribution is the distribution of the ratio of two independent uniform random variables. Specifically, suppose \(U \in [-1,1]\) and \(V \in [0,1]\) are independent and uniformly distributed. Then \(R = U/V\) has the ratio of uniforms distribution. The plot below shows a histogram based on 10,000 samples from the ratio of uniforms distribution2.

    The histogram has a flat section in the middle and then curves down on either side. This distinctive shape is called a “table mountain”. The density of \(R\) also has a table mountain shape.

    And here is the density plotted on top of the histogram.

    A formula for the density of \(R\) is

    \(\displaystyle{h(R) = \begin{cases} \frac{1}{4} & \text{if } -1 \le R \le 1, \\\frac{1}{4R^2} & \text{if } R < -1 \text{ or } R > 1.\end{cases}}\)

    The first case in the definition of \(h\) corresponds to the flat part of the table mountain. The second case corresponds to the sloping curves. The rest of this post use geometry to derive the above formula for \(h(R)\).

    Calculating the density

    The point \((U,V)\) is uniformly distributed in the box \(B=[-1,1] \times [0,1]\). The image below shows an example of a point \((U,V)\) inside the box \(B\).

    We can compute the ratio \(R = U/V\) geometrically. First we draw a straight line that starts at \((0,0)\) and goes through \((U,V)\). This line will hit the horizontal line \(y=1\). The \(x\) coordinate at this point is exactly \(R=U/V\).

    In the above picture, all of the points on the dashed line map to the same value of \(R\). We can compute the density of \(R\) by computing an area. The probability that \(R\) is in a small interval \([R,R+dR]\) is

    \(\displaystyle{\frac{\text{Area}(\{(u,v) \in B : u/v \in [R, R+dR]\})}{\text{Area}(B)} = \frac{1}{2}\text{Area}(\{(u,v) \in B : u/v \in [R, R+dR]\}).}\)

    If we can compute the above area, then we will know the density of \(R\) because by definition

    \(\displaystyle{h(R) = \lim_{dR \to 0} \frac{1}{2dR}\text{Area}(\{(u,v) \in B : u/v \in [R, R+dR]\})}.\)

    We will first work on the case when \(R\) is between \(-1\) and \(1\). In this case, the set \(\{(u,v) \in B : u/v \in [R, R+dR]\}\) is a triangle. This triangle is drawn in blue below.

    The horizontal edge of this triangle has length \(dR\). The perpendicular height of the triangle from the horizontal edge is \(1\). This means that

    \( \displaystyle{\text{Area}(\{(u,v) \in B : u/v \in [R, R+dR]\}) =\frac{1}{2}\times dR \times 1=\frac{dR}{2}}.\)

    And so, when \(R \in [-1,1]\) we have

    \(\displaystyle{h(R) = \lim_{dR\to 0} \frac{1}{2dR}\times \frac{dR}{2}=\frac{1}{4}}.\)

    Now let’s work on the case when \(R\) is bigger than \(1\) or less than \(-1\). In this case, the set \(\{(u,v) \in B : u/v \in [R, R+dR]\}\) is again triangle. But now the triangle has a vertical edge and is much skinnier. Below the triangle is drawn in red. Note that only points inside the box \(B\) are coloured in.

    The vertical edge of the triangle has length \(\frac{1}{R} – \frac{1}{R+dR}= \frac{dR}{R(R+dR)}\). The perpendicular height of the triangle from the vertical edge is \(1\). Putting this together

    \( \displaystyle{\text{Area}(\{(u,v) \in B : u/v \in [R, R+dR]\}) =\frac{1}{2}\times \frac{dR}{R(R+dR)} \times 1=\frac{dR}{2R(R+dR)}}.\)

    And so

    \(\displaystyle{h(R) = \lim_{dR \to 0} \frac{1}{2dR} \times \frac{dR}{2 R(R+dR)} = \frac{1}{4R^2}}.\)

    And so putting everything together

    \(\displaystyle{h(R) = \begin{cases} \frac{1}{4} & \text{if } -1 \le R \le 1, \\\frac{1}{4R^2} & \text{if } R < -1 \text{ or } R > 1.\end{cases}}\)

    Footnotes and references

    1. https://ieeexplore.ieee.org/document/718718 ↩︎
    2. For visual purposes, I restricted the sample to values of \(R\) between \(-8\) and \(8\). This is because the ratio of uniform distribution has heavy tails. This meant that there were some very large values of \(R\) that made the plot hard to see. ↩︎

  • The discrete arcsine distribution

    The discrete arcsine distribution is a probability distribution on \(\{0,1,\ldots,n\}\). It is a u-shaped distribution. There are peaks at \(0\) and \(n\) and a dip in the middle. The figure below shows the probability distribution function for \(n=10,15, 20\).

    The probability distribution function of the arcsine distribution is given by

    \(\displaystyle{p_n(k) = \frac{1}{2^{2n}}\binom{2k}{k}\binom{2n-2k)}{n-k}\text{ for } k \in \{0,1,\ldots,n\}}\)

    The discrete arcsine distribution is related to simple random walks and to an interesting Markov chain called the Burnside process. The connection with simple random walks is explained in Chapter 3, Volume 1 of An Introduction to Probability and its applications by William Feller. The connection to the Burnside process was discovered by Persi Diaconis in Analysis of a Bose-Einstein Markov Chain.

    The discrete arcsine distribution gets its name from the continuous arcsine distribution. Suppose \(X_n\) is distributed according to the discrete arcsine distribution with parameter \(n\). Then the normalized random variables \(X_n/n\) converges in distribution to the continuous arcsine distribution on \([0,1]\). The continuous arcsine distribution has the probability density function

    \(\displaystyle{f(x) = \frac{1}{\pi\sqrt{x(1-x)}} \text{ for } 0 \le x \le 1}\)

    This means that continuous arcsine distribution is a beta distribution with \(\alpha=\beta=1/2\). It is called the arcsine distribution because the cumulative distribution function involves the arcsine function

    \(\displaystyle{F(x) = \int_0^x f(y)dy = \frac{2}{\pi}\arcsin(\sqrt{x}) \text{ for } 0 \le x \le 1}\)

    There is another connection between the discrete and continuous arcsine distributions. The continuous arcsine distribution can be used to sample the discrete arcsine distribution. The two step procedure below produces a sample from the discrete arcsine distribution with parameter \(n\):

    1. Sample \(p\) from the continuous arcsine distribution.
    2. Sample \(X\) from the binomial distribution with parameters \(n\) and \(p\).

    This means that the discrete arcsine distribution is actually the beta-binomial distribution with parameters \(\alpha = \beta =1/2\). I was surprised when I was told this, and couldn’t find a reference. The rest of this blog post proves that the discrete arcsine distribution is an instance of the beta-binomial distribution.

    As I showed in this post, the beta-binomial distribution has probability distribution function:

    \(\displaystyle{q_{\alpha,\beta,n}(k) = \binom{n}{k}\frac{B(k+\alpha, n-k+\alpha)}{B(a,b)}},\)

    where \(B(x,y)=\frac{\Gamma(x)\Gamma(y)}{\Gamma(x+y)}\) is the Beta-function. To show that the discrete arc sine distribution is an instance of the beta-binomial distribution we need that \(p_n(k)=q_{1/2,1/2,n}(k)\). That is

    \(\displaystyle{ \binom{n}{k}\frac{B(k+1/2, n-k+1/2)}{B(1/2,1/2)} = \frac{1}{2^{2n}}\binom{2k}{k}\binom{2n-2k}{n-k}}\),

    for all \(k = 0,1,\ldots,n\). To prove the above equation, we can first do some simplifying to \(q_{1/2,1/2,n}(k)\). By definition

    \(\displaystyle{\frac{B(k+1/2, n-k+1/2)}{B(1/2,1/2)} = \frac{\frac{\Gamma(k+1/2)\Gamma(n-k+1/2)}{\Gamma(n+1)}}{\frac{\Gamma(1/2)\Gamma(1/2)}{\Gamma(1)}} = \frac{1}{n!}\frac{\Gamma(k+1/2)}{\Gamma(1/2)}\frac{\Gamma(n-k+1/2)}{\Gamma(1/2)}}\),

    where I have used that \(\Gamma(m)=(m-1)!\) factorial if \(m\) is a natural number. The Gamma function \(\Gamma(x)\) also satisfies the property \(\Gamma(x+1)=x\Gamma(x)\). Using this repeatedly gives

    \(\displaystyle{\Gamma(k+1/2) = (k-1/2) \times (k-3/2) \times \cdots \times \frac{3}{2}\times\frac{1}{2}\times\Gamma(1/2) }. \)

    This means that

    \(\displaystyle{\frac{\Gamma(k+1/2)}{\Gamma(1/2)} = (k-1/2) \times (k-3/2) \times \cdots \times \frac{3}{2}\times\frac{1}{2} = \frac{(2k-1)\times(2k-3)\times \cdots \times 3 \times 1}{2^k}=\frac{(2k-1)!!}{2^k}},\)

    where \((2k-1)!!=(2k-1)\times (2k-3)\times\cdots \times 3 \times 1\) is the double factorial. The same reasoning gives

    \(\displaystyle{\frac{\Gamma(n-k+1/2)}{\Gamma(1/2)} =\frac{(2n-2k-1)!!}{2^{n-k}}}.\)

    And so

    \(\displaystyle{q_{1/2,1/2,n}(k) =\frac{1}{2^nk!(n-k)!}(2k-1)!!(2n-2k-1)!!}.\)

    We’ll now show that \(p_n(k)\) is also equal to the above final expression. Recall

    \(\displaystyle{p_n(k) = \frac{1}{2^{2n}} \binom{2k}{k}\binom{2(n-k)}{n-k} = \frac{1}{2^{2n}}\frac{(2k)!(2(n-k))!}{k!k!(n-k)!(n-k)!} = \frac{1}{2^nk!(n-k)!}\frac{(2k)!}{k!2^k}\frac{(2n-2k)!}{(n-k)!2^{n-k}}}.\)

    And so it suffices to show \(\frac{(2k)!}{k!2^k} = (2k-1)!!\) (and hence \(\frac{(2n-2k)!}{(n-k)!2^{n-k}}=(2n-2k-1)!!\)). To see why this last claim holds, note that

    \(\displaystyle{\frac{(2k)!}{k!2^k} = \frac{(2k)\times (2k-1)\times(2k-2)\times\cdots\times 3 \times 2 \times 1}{(2k)\times (2k-2)\times \cdots \times 2} = (2k-1)!!}\)

    Showing that \(p_{n}(k)=q_{n,1/2,1/2}(k)\) as claimed.

  • The sample size required for importance sampling

    My last post was about using importance sampling to estimate the volume of high-dimensional ball. The two figures below compare plain Monte Carlo to using importance sampling with a Gaussian proposal. Both plots use \(M=1,000\) samples to estimate \(v_n\), the volume of an \(n\)-dimensional ball

    A friend of mine pointed out that the relative error does not seem to increase with the dimension \(n\). He thought it was too good to be true. It turns out he was right and the relative error does increase with dimension but it increases very slowly. To estimate \(v_n\) the number of samples needs to grow on the order of \(\sqrt{n}\).

    To prove this, we will use the paper The sample size required for importance sampling by Chatterjee and Diaconis [1]. This paper shows that the sample size for importance sampling is determined by the Kullback-Liebler divergence. The relevant result from their paper is Theorem 1.3. This theorem is about the relative error in using importance sampling to estimate a probability.

    In our setting the proposal distribution is \(Q=\mathcal{N}(0,\frac{1}{n}I_n)\). That is the distribution \(Q\) is an \(n\)-dimensional Gaussian vector with mean \(0\) and covariance \(\frac{1}{n}I_n\). The conditional target distribution is \(P\) the uniform distribution on the \(n\) dimensional ball. Theorem 1.3 in [1] tells us how many samples are needed to estimate \(v_n\). Informally, the required sample size is \(M = O(\exp(D(P \Vert Q)))\). Here \(D(P\Vert Q)\) is the Kullback-Liebler divergence between \(P\) and \(Q\).

    To use this theorem we need to compute \(D(P \Vert Q)\). Kullback-Liebler divergence is defined as integral. Specifically

    \(\displaystyle{D(P\Vert Q) = \int_{\mathbb{R}^n} \log\frac{P(x)}{Q(x)}P(x)dx}\)

    Computing the high-dimensional integral above looks challenging. Fortunately, it can reduced to a one-dimensional integral. This is because both the distributions \(P\) and \(Q\) are rotationally symmetric. To use this, define \(P_r,Q_r\) to be the distribution of the norm squared under \(P\) and \(Q\). That is if \(X \sim P\), then \(\Vert X \Vert_2^2 \sim P_R\) and likewise for \(Q_R\). By the rotational symmetry of \(P\) and \(Q\) we have

    \(D(P\Vert Q) = D(P_R \Vert Q_R).\)

    We can work out both \(P_R\) and \(Q_R\). The distribution \(P\) is the uniform distribution on the \(n\)-dimensional ball. And so for \(X \sim P\) and any \(r \in [0,1]\)

    \(\mathbb{P}(\Vert X \Vert_2^2 \le r) = \frac{v_n r^n}{v_n} = r^n.\)

    Which implies that \(P_R\) has density \(P_R(r)=nr^{n-1}\). This means that \(P_R\) is a Beta distribution with parameters \(\alpha = n, \beta = 1\). The distribution \(Q\) is a multivariate Gaussian distribution with mean \(0\) and variance \(\frac{1}{n}I_n\). This means that if \(X \sim Q\), then \(\Vert X \Vert_2^2 = \sum_{i=1}^n X_i^2\) is a scaled chi-squared variable. The shape parameter of \(Q_R\) is \(n\) and scale parameter is \(1/n\). The density for \(Q_R\) is therefor

    \(Q_R(r) = \frac{n^{n/2}}{2^{n/2}\Gamma(n/2)}r^{n/2-1}e^{-nx/2}\)

    The Kullback-Leibler divergence between \(P\) and \(Q\) is therefor

    \(\displaystyle{D(P\Vert Q)=D(P_R\Vert Q_R) = \int_0^1 \log \frac{P_R(r)}{Q_R(r)} P_R(r)dr}\)

    Getting Mathematica to do the above integral gives

    \(D(P \Vert Q) = -\frac{1+2n}{2+2n} + \frac{n}{2}\log(2 e) – (1-\frac{n}{2})\log n + \log \Gamma(\frac{n}{2}).\)

    Using the approximation \(\log \Gamma(z) \approx (z-\frac{1}{2})\log(z)-z+O(1)\) we get that for large \(n\)

    \(D(P \Vert Q) = \frac{1}{2}\log n + O(1)\).

    And so the required number of samples is \(O(\exp(D(P \Vert Q)) = O(\sqrt{n}).\)

    [1] Chatterjee, Sourav, and Persi Diaconis. “THE SAMPLE SIZE REQUIRED IN IMPORTANCE SAMPLING.” The Annals of Applied Probability 28, no. 2 (2018): 1099–1135. https://www.jstor.org/stable/26542331. (Public preprint here https://arxiv.org/abs/1511.01437)