# Upper bounds on correlations and Cronbach’s alpha for discrete random variables

This is a small tutorial on how to either use the R code (if you are an R user) or the web application (if you prefer a user-friendly GUI) that will calculate the upper bound on (positive) correlations between random variables with an arbitrary number of categories.

The first thing to know is that your categories should start at 0 (not 1). If you are using the R code, please make sure your categories start there. If you are using the GUI, it will check (variable by variable) whether or not the categories start at 0 and will make the necessary adjustments for you.

The overall interface is very simple:

The first step will be to upload your data where it says “Browse”. Notice that:

(1) This app ONLY takes .csv files. If you are an SPSS user and are unfamiliar with how to save your SPSS (extension .sav) data as a .csv file, here is an easy explanation. Trust me, it’s A LOT easier than you think.

(2) It will use ALL the variables in your data file. If you have other variables or covariates which you are not interested in correlating, you have to remove them before uploading the data.

(3) It doesn’t handle missing data. You need to provide a .csv file with no missing elements so you need to ‘pre-process’ it before you upload it.

(4) It can only handle numbers. If you leave your variable names, it will give you an error. So make sure those are removed as well before uploading anything.

And…. that’s pretty much it. Once you’ve pre-processed the data and uploaded the file just click “Run” and it should give you something that looks like this:

The interpretation is very simple. It uses the number of categories and the probabilities of endorsing each category to calculate the maximum possible correlations that you could have observed. Then it uses this convenient formula that operates exclusively on the correlation (or the covariance) matrix to calculate Cronbach’s alpha. Keep in mind that this would be the standardized Cronbach’s alpha (as opposed to the raw one). But you can always use it as a point of reference. Or work backwards from the maximum possible correlation matrix and “unstandardize” it to obtain the maximum possible covariance matrix.

If you have many variables, the correlation matrix won’t be displayed. But you can download it as a .csv file for inspection or click on ‘copy’ to copy-paste it in Microsoft Words or some text editor of your choice. Cronbach’s alpha is always calculated.

And… that is pretty much it.

And if you are an R user and feel more at home with code, HERE IS THE CODE THAT RUNS THE ANALYSIS.

# Is there life beyond the 4th (central) moment?

Us, academic reviewers, can be very finicky creatures (BTW don’t @me because deep down you know it is true. We’ve all been Reviewer 2 at least once in our careers 😜). Sometimes we miss the author(s)’ key points. Sometimes we review the paper that we think was written as opposed to the one that was actually written. And sometimes, we (serendipitously) offer the kind of keen insight into a problem that makes an author drop everything they’re doing and think “well, well, well… now THIS is interesting”.  This story is about the latter type of review.

I will omit a lot of details here because our article has just been accepted (yay!) and we gotta respect the process. But the really cool stuff happened after the review so we’re all good. Anyway, here it goes: Reviews come in and they’re mostly short and positive. Change a few things here, add a few things there and we’re good to go. So everything is very hunky dory. As we’re working through them, this fine little piece catches my eyes (redacted):

It’s interesting to note that _________________  for uniform distributions (negative kurtosis), whereas _________________ concluded that _______________ is especially not robust to non-normal distributions with excess kurtosis (i.e., overestimated with negative kurtosis and underestimated with positive kurtosis), which does not seem to be in line with findings from this study. I wonder what the authors’ insight is on this particular aspect.

Now, it just so happens that _________________ was very thorough with their description of their simulation study and I could actually reproduce what they were doing. Also, as luck would have it, _________________ used a data-generating method that I am very, VERY familiar with. They used the 5th order power polynomial transformation which I myself used in the very first article I ever published. And here is the thing. If you compare the data _________________used as opposed to the one I used (uniform distributions) They appear to be very, very similar if you just look at the descriptives. Here’s the R output comparing the uniform distribution to the distribution used by _________________ (results are standardized to ease comparison):

vars     n mean sd median trimmed   mad    min  max range skew kurtosis  se them 1000000   0   1     0        0 1.41 -2.34 2.34  4.68     0    -1.4   0 me   1000000   0   1     0        0 1.28 -1.73 1.73  3.46     0    -1.2   0

So… at least on paper these two things look the same. And I’d have a hard time believing that a -0.2 difference between (excess) kurtosis would result in all of this. Nevertheless, the advantage of working with the 5th order polynomial approach is that (whether you are interested in them or not), it allows the user control over the 5th and 6th central moments (so the next 2 moments after kurtosis).  I would venture out a guess as to say you have probably never heard of those. Maybe you have never even seen how they look like. Luckily, with a little bit of algebra (and some help from Kendall & Stuart (1977) who defined the first six cumulants) it’s not too hard to show that the 5th and 6th (standardized) central moments are:

$\gamma_5=\frac{\int_{-\infty}^{\infty}(x-\mu)^5dF(x)}{[\int_{-\infty}^{\infty}(x-\mu)^2dF(x)]^{5/2}}-10\gamma_3$

$\gamma_6=\frac{\int_{-\infty}^{\infty}(x-\mu)^6dF(x)}{[\int_{-\infty}^{\infty}(x-\mu)^2dF(x)]^{3}}-15\gamma_4-10\gamma_3-15$

where $F(x)$ is the cumulative density function (CDF), $\gamma_3$ is the skewness and $\gamma_4$ is the excess kurtosis. Now I know that the distributions that _________________ used were symmetric along the mean so $\gamma_5=0$ because all the odd moments of a symmetric distribution are 0. And I know that $\gamma_6=22$ because if you know the coefficients used in the 5th order transformation, you can sort of  “reverse-engineer” what the central moments of the distribution were. Which means all I had to do was figure out what $(\gamma_5, \gamma_6)$ are for the uniform distribution. Again, for the uniform distribution $\gamma_5=0$ by the same symmetry argument. So I just need to know that $\gamma_6$ is. Using the standard uniform CDF (so defined in the interval [0,1]) and the definition of central moments I need to solve for:

$\gamma_6=\frac{\int_{0}^{1}(x-1/2)^6dx}{[\int_{0}^{1}(x-1/2)^2dx]^{3}}-15(-6/5)-10(0)-15=\frac{48}{7} \approx 6.9$

So… it seems like I can more or less match these two distributions all the way up to the 6th moment, which is where they become VERY different. Now, as expected, the big reveal is…. how do they look like when you graph them? Well, they look like this:

Yup, yup… there seems to be some bimodality going on here. It sort of looks like a beta distribution with equal parameters of 0.5, with the exception that the beta distribution is not defined outside the [0,1] range.

As usual, the next step is to figure out whether using one VS the other one could matter. I happen to be doing a little bit of research into tests for differences between correlations so I thought this would be a good opportunity to try this one out. I’m starting off with the easy-cheezy, t-test for two dependent correlations that Hotelling (1940) came up with. In any case, the t-statistic is expressed as follows:

$t=\frac{(r_{x_{1}y}-r_{x_{2}y})\sqrt{(n-3)(1+r_{x_{1}x_{2}})}}{det(\mathbf{R})}$

where $r_{x_{1}y}$ and $r_{x_{2}y}$ are the two correlations whose difference you are interested in, $r_{x_{1}x_{2}}$ is the correlation that induces the dependency and $det(\mathbf{R})$ is the determinant of the correlation matrix. So you have three variables at play here, $x_1, x_2$ and $y$ so the correlation matrix has dimensions $3 \times 3$.

Alrighty, so here is the set-up for the small-scale simulation. I am gonna check the empirical Type I error rate of this t-test for the differences between correlations. I am setting up the 3 variables to be correlated equally with $\rho=0.5$ in the population. As simulation “conditions” I am using the two types of non-normalities discussed above: the uniform distribution VS the 5th-order-polynomial-transformed distribution. BTW, I am going to start calling them “Headrick distributions” in honour of Dr. Todd Headrick who laid the theoretical foundation for this area of research).

The uniform distribution part is easy because we can just use a Gaussian copula which ensures the distribution of the marginals is in fact the distribution that we intend (uniform).  The 5th order polynomial part is a little bit more complicated. But since I figured out how to do this in R a while ago, then I know what should go in place of the intermediate correlation matrix (if this concept is new to you, please check out this blog of mine). In any case, simulation code:

set.seed(112) library(MASS)

nn<-100 #sample size reps<-1000 #replications h<-double(reps) #vector to store results c<-double(reps) #vector to store results

for(i in 1:reps){

##5th order polynomial transform

Z<-mvrnorm(nn, c(0,0,0), matrix(c(1,.60655,.60655,.60655,1,.60655,.60655,.60655,1),3,3)) z1<-Z[,1] z2<-Z[,2] z3<-Z[,3]

y1<-1.643377*z1-0.319988*z1^3+0.011344*z1^5 y2<-1.643377*z2-0.319988*z2^3+0.011344*z2^5 y3<-1.643377*z3-0.319988*z3^3+0.011344*z3^5 
Y<-cbind(y1,y2,y3)

############################################## 
##Gaussian copula

X <- mvrnorm(nn, c(0,0,0), matrix(c(1,.5189,.5189,.5189,1,.5189,.5189,.5189,1),3,3)) U <- pnorm(X)

q1<-qunif(U[,1] ,min=0, max=1) q2<-qunif(U[,2] ,min=0, max=1) q3<-qunif(U[,3] ,min=0, max=1)

Q<- cbind(q1,q2,q3)

###############################################

r12<-cor(y1,y2) r13<-cor(y1,y3) r23<-cor(y2,y3)

rq1q2 <- cor(q1, q2) rq1q3 <- cor(q1, q3) rq2q3 <- cor(q2, q3)

################################################ ##Test statistic from Hotelling (1940)

tt1<-(rq1q2-rq1q3)*sqrt(((nn-3)*(1+rq2q3)) / (2*det(cor(Q)))) c[i]<-2*pt(-abs(tt1),df=nn-3)

tt2<-(r12-r13)*sqrt(((nn-3)*(1+r12)) / (2*det(cor(Y)))) h[i]<-2*pt(-abs(tt2),df=nn-3)

} ############################################### #Empirical Type I error rates

 >sum(h<.05)/reps
[1] 0.09
 >sum(c<.05)/reps
[1] 0.056

Well, well, well… it appears that there *is* something interesting about the Headrick distribution that isn’t just there for the uniform. I mean, the uniform distribution is showing me a very good empirical Type I error rate. I’m sure a few more replications would get that rid of that additional .006 . But for the Headrick distribution the Type I error rate almost doubled. It does seem like whatever weirdness is going on in those tails may be playing a role on this… or maybe not? You see, I am beginning to wonder these days whether or not the definitions of moments beyond the variance are actually all that useful. I myself have a small collection of pictures of distributions that look somewhat (or VERY) different, yet have the same mean, variance, skewness and excess kurtosis. I use them in my intro classes to highlight to my students the importance of plotting your data.

Now, it could very well be the case that moments beyond the 4th only matter to differences of correlations and nothing else. Or maybe they do matter to other techniques. Which makes me wonder further… how far down the line of moments are we supposed to go before we can safely claim that X or Y statistical technique is robust? Sometimes I think we would be better off trying to make claims about families of probability distributions as opposed to distributions that exhibit certain values of certain moments. After all, Stoyanov’s famous book on counter examples in probability has a whole section of  “moment problems” that highlights how one can build families of distributions with the same (finite) number of moments yet they are not the same type of distributions.

In any case, I guess the only sensible conclusion I can get from this is the good, ol’ “clearly, more research is needed” ¯\_(ツ)_/¯

# How to simulate correlated log-normal random variables THE RIGHT WAY

This came out of an email exchange that I had with my dear friend Ben Shear and I eventually realized it could benefit more people. If you have two log-normal random variables how can you correlate them the right way? When I say the right way I mean that you both get the population correlation that you want and the 1-dimensional marginals that you want.

Naive approach #1 (Wrong): Correlate first, transform later

This one is a popular one:

library(MASS)
mult_norm<- mvrnorm(1000, c(0,0), matrix(c(1,.5,.5,1),2,2))
log_norm<- exp(mult_norm)

This one gets you the right marginals (they are, indeed, log-normal). But the wrong correlation matrix as you can see:

cor(mult_norm) [,1]      [,2] [1,] 1.0000000 0.4967083 [2,] 0.4967083 1.0000000

 > cor(log_norm) [,1]      [,2] [1,] 1.0000000 0.4419539 [2,] 0.4419539 1.0000000

The resulting correlation matrix is biased downwards. So that won’t do.

Naive approach #2 (Wrong as well): Transform first, correlate later

This one is also very appealing because of its simplicity and because it relies on the construction of a bivariate normal distribution. Like, if $(X,Z)$ are each independent and identically distributed $\mathcal{N} \sim (0,1)$ define $Y= X\rho + Z\sqrt{1-\rho^{2}}$. Then $(X,Y)$ are bivariate normally distributed with $\mu_{X}=\mu_{Y}=0, \sigma_{X}=\sigma_{Y}=1$ and population correlation coefficient $\rho$.

What would happen then if we did $A=e^{X}, B=e^{Z}$ and did $Y= A\rho + B\sqrt{1-\rho^{2}}$?

In this case, $A$ would still be log-normal and $cor(A,Y)=\rho$. But $Y$ would NOT be log-normally distributed. The Dufresne (2009) paper is the standard reference for this, but there have been some interesting developments, like this approximation using the log-skew-normal distribution. In any case, the distribution of $Y$ isn’t known.

A third (correct) way

This is essentially a re-hash of Appendix A in Astivia & Zumbo (2017) where it’s not too difficult to show that if $(X,Y)$ are standard bivariate normal with correlation coefficient $\rho$ then the correlation between $A=e^{X}, B=e^{Y}$ is:

$\displaystyle cor(A,B)=\frac{e^{\rho}-1}{e-1}$

So what we can do is solve for $\rho$ above and find that:

$\displaystyle \rho=ln(cor(A,B)(e-1)+1)$

And what we do is that we go back to Naive Approach #1 and plug **that** as the initial correlation coefficient so that, through the process of exponentiation, we get back the value for $cor(A,B)$ that we want our log-normal marginal distributions to have in the population.

Small simulation check:

library(MASS)

###I want a population correlation between ###the log-normals of 0.5 so:
rr<-double(1000)

for(i in 1:1000){ mult_norm<-mvrnorm(1000, c(0,0), matrix(c(1,log(.5*(exp(1)-1)+1),log(.5*(exp(1)-1)+1),1),2,2))

log_norm<-exp(mult_norm)

rr[i]<-cor(log_norm)[1,2]
} mean(rr) >0.5019529

And the joint distribution would look like this:

We usually call the correlation that we need to set for the initial bivariate normal the “intermediate correlation” and it gives us some interesting insights into how the correlation coefficient behaves when the marginals are not normal.

For example here if I wanted to give my log-normals a correlation $\rho=-0.5$ I’d do:

 log(-.5*(exp(1)-1)+1) [1] -1.959995

And find out that I can’t. Two log-normal random variables generated from a standard bivariate normal distribution CANNOT be correlated at -0.5. The maximum possible negative correlation is actually $-1/e$.

# Multivariate Poisson & Multivariate Exponential Distributions (not everything needs a copula ;)

While I am preparing for a more “in-depth” treatment of this Twitter thread that sparked some interest (thank my lucky stars!) , I ran into a couple of curious distributions that I think highlight an important lesson: not EVERYTHING needs a copula 😉 Sure, they’re flexible and powerful modelling approaches. But at the end of the day there are other methods to create multivariate, non-normal distributions. And many of them have very elegant theoretical properties that allow us to expand our the intuitions about these distributions from the univariate to the multivariate setting.

So here I present two distributions which can be generalized from their univariate to a multivariate definition without invoking a copula.

Multivariate Poisson Distribution

The Poisson distribution is closed under convolutions. For purposes of this post, that means that if $X_{1}$ and $X_{2}$ are independent, Poisson-distributed (with parameters $\lambda_{1}, \lambda_{2}$ respectively) then $Y=X_{1}+X_{2}$ is also Poisson-distributed, (with parameter… Yup! You got it! $\lambda_{Y}=\lambda_{1}+\lambda_{2}$).

So let’s start easy, the bivariate case. To construct a bivariate Poisson random vector $(X,Y)'$ we can use the following stochastic representation:

$X = X_{1} + X_{1,2}$

$Y = X_{2} + X_{1,2}$

Where $X_{1}, X_{2}, X_{1,2}$ are independent Poisson random variables with parameters $\lambda_{1}, \lambda_{2}, \lambda_{1,2}$ respectively. Since Poisson distributions are closed under convolutions, $X$ and $Y$ are Poisson distributed with variance $\lambda_{1}+\lambda_{1,2}, \lambda_{2}+\lambda_{1,2}$ respectively, and covariance $Cov(X, Y) = \lambda_{1,2}$. Notice that this construction implies the restriction $Cov(X, Y) \geq 0$.

Using the above property we can derive the joint probability function of $(X,Y)'$. Consider $P(X=r, Y=s)$ fixed values $r,s$ Since $X, Y, X_{12}$ can take any values between 0 and $min(r,s)$ and $X_{1}, X_{2}, X_{1,2}$ are mutually independent then we can use this property to define the joint probability function as:

$P(X=r, Y=s)=\sum_{k=0}^{min(r,s)}f_1(r-k)f_2(s-k)f_{1,2}(k)$

where $f_1, f_2, f_{1,2}$ are the probability functions of $X_1, X_2, X_{1,2}$ respectively. By making the proper substitutions in the $f's$ and some collecting of terms we have:

$P(X=r, Y=s)=e^{(-\lambda_1-\lambda_2-\lambda_{1,2})}\frac{\lambda^{r}_{1}}{r!}\frac{\lambda^{s}_{2}}{s!}\sum_{k=0}^{min(r,s)}(r-k)(s-k)k!\left ( \frac{\lambda_{1,2}}{\lambda_1\lambda_2} \right ) ^{k}$

From this process I could expand it to, say, a trivariate Poisson random variable by expressing the 3-D vector $(X, Y, Z)'$ as:

$X = X_{1} + X_{1,2} + X_{1,3} +X_{1,2,3}$

$Y = X_{2} + X_{1,2} + X_{2,3} +X_{1,2,3}$

$Z = X_{3} + X_{1,3} + X_{2,3} +X_{1,2,3}$

Where all the X’s are themselves independent, Poisson distributed and the terms with double (and triple) subscript would control the level of covariance among the Poisson marginal distributions.  If you repeat this iteratively adding more and more terms to the summation then you can increase the dimensions of the multivariate Poisson distribution.

Multivariate Exponential Distribution

The univariate exponential distribution is also (sort of) closed under convolution. But, in this very specific case, it’s closed under weighted minima convolution. For this post, that means that if $Y_1, Y_2, Y_3, ..., Y_n$ are independent, exponential random variables, then $min(w_1Y_1, w_2Y_2, w_3Y_3, ..., w_nY_n)$ is also exponentially-distributed for $w_1, w_2, w_3, ..., w_n >0$.

We can start very similarly as with the previous case by defining how the bivariate distribution would look like. Here we start with:

$X = Y_{1} + Y_{1,2}$

$Y = Y_{2} + Y_{1,2}$

where $Y_1, Y_2, Y_{1,2}$ are independent, exponentially-distributed random variables with parameters $\lambda_{1}, \lambda_{2}, \lambda_{1,2}$ (I SWEAR to God the standard mathematical notation for the parameter of both the exponential and the Poisson distribution is $\lambda$. Of course $Y_1, Y_2, Y_{1,2}$ would be chosen as the minima of their respective sequences of exponential random variables. Because of this $X, Y$ are also exponentially-distributed with parameters $\lambda_{1}+\lambda_{1,2}, \lambda_{2}+\lambda_{1,2}$ respectively. Since $Y_1, Y_2, Y_{1,2}$ are independent, then we have:

$P(X>x, Y>y)=P(Y_1>x, Y_{1,2}>x, Y_2>y, Y_{1,2}>y)=$

$e^{ \lambda_1x}e^{\lambda_2y}e^{\lambda_{1,2}max(x,y)}$

And the joint cumulative density function of the bivariate vector $(X,Y)'$ would then be:

$F(y_1, y_2)=(1-e^{-\lambda_{1}x})(1-e^{-\lambda_{2}y})(1-e^{-\lambda_{1,2}max(x,y)})$

Moral of the story

If you know me, you’ll know that I tend to be overly critical of the things I like the most, which is a habit that doesn’t always makes me a lot of friends, heh. Because I like copula modelling and I like the idea of non-normal, multivariate structures, I also like to see and understand the cases where defining multivariate structures that do not need a copula may give us insights. For the particular two cases above, I am exploiting the fact that sums of these types of random variables also result in the same type of random variable (i.e., closed under convolution) which, for better or worse, is a very useful property that not many univariate probability distributions have. Nevertheless, when they *do* have it, it is perhaps wise to use them because, at the end of the day, using copulas to either simulate or model multivariate data does not imply the copula distribution *becomes* the “multivariate” version of that distribution. A Gaussian copula with gamma-distributed marginals is not a “multivariate gamma distribution”. It is… well… a Gaussian copula with gamma distributed marginals. If you change the copula function for something else (say an Archimidean copula) the multivariate properties change as well. Which brings us to a very sobering realization: with the exception of some very select types of multivariate distributions (usually those closed under convolution) we don’t always have well-defined extensions of multivariate distributions. For example, there are more than 10 different ways to define distributions that would satisfy what one would call a “multivariate t distribution”. Whichever characterization one chooses is usually contingent on the intended use for it. I guess this helps highlight one of the oldest pieces of advice I ever received: You know you have become a good methodologist when you realize the only correct answer to every data analysis question is simply “it depends”.

# Power analysis for multiple regression with non-normal data

This app will perform computer simulations to estimate the power of the t-tests within a multiple regression context under the assumption that the predictors and the criterion variable are continuous and either normally or non-normally distributed. If you would like to see why I think this is important to keep in mind, please read this blog post.

When you first click on the app it looks like this:

What *you*, the user, needs to provide it with is the following:

The number of predictors. It can handle anything from 3 to 6 predictors. When you have more than that the overall aesthetics of the app is simply too crowded.  The default is 3 predictors.

The regression coefficients (i.e, the standardized effect sizes) that hold in the population. The default is 0.3. The app names them “x1, x2, x3,…x6”

The skewness and excess kurtosis of the data for each predictor AND for the dependent variable (the app calls it “y”). Please keep on reading to see how you should choose those. The defaults at this point are skewness of 2 and an excess kurtosis of 7.

The pairwise correlations among the predictors. I think this is quite important because the correlation among the predictors plays a role in calculating the standard error of the regression coefficients. So you can either be VERY optimistic and place those at 0 (predictors are perfectly orthogonal with one another) OR you can be very pessimistic and give them a high correlation (multicollinearity). The default inter-predictor correlation is 0.5.

The sample size. The default is 200.

The number of replications for the simulation. The default is 100.

Now, what’s the deal with the skewness and excess kurtosis? A lot of people do not know this but you cannot go around choosing values of skewness and excess kurtosis all willy-nilly. There is a quadratic relationship between the possible values of skewness and excess kurtosis that specifies they MUST be chosen according to the inequality kurtosis >skewness^2-2 . If you don’t do this, it will spit out an error. Now, I am **not** a super fan of the algorithm needed to generate data with those population-specified values of skewness and excess kurtosis. For many AND even more reasons. HOWEVER, I needed to choose something that was both sufficiently straight-forward to implement and not very computationally-intensive, so the 3rd order polynomial approach will have to do.

Now, the exact boundaries of what this method can calculate are actually smaller than the theoretical parabola. However, for practical purposes, as long as you choose values of kurtosis which are sufficiently far apart from the square of the skewness, you should be fine. So, a combo like skewness=3, kurtosis=7 would give it trouble. But something like skewness=3, kurtosis=15 would be perfectly fine.

A hypothetical run would look like this:

So the output under Results is the empirical, simulated power for each regression coefficient at the sample size selected. In this case, they gravitate around 60%.

Oh! And if for whatever reason you would like to have all your predictors be normal, you can set the values of skewness and kurtosis to 0. In fact, in that situation you would end up working with a multivariate normal distribution.

# Power Analysis for the Pearson correlation with non-normal data

This app will perform computer simulations to estimate the power of the t-test for the Pearson bivariate correlation under the assumption that the data are continuous and either bivariate normally distributed or non-normally distributed. To see why this is important, please check out this blog post.

It is very straightforward to use. When you click on the app, it will look like this:

What *you*, the user, needs to provide it with is the following:

The population correlation (i.e., the effect size) for which you would like to obtain power. The default is 0.3

The type of distributions you would like to correlate together. Right now it can handle the chi-square distribution (where skewness is controlled through the degrees of freedom), the uniform distribution (to have a symmetric distribution with negative kurtosis) and the binomial distribution where one can control the number of response categories (size) and the probability parameter. This will soon be replaced by a multinomial distribution so that the probability of every marginal response option can be specified.

The sample size. The default is 20

The number of replications for the simulation. The default is 100.

Now, what is the app actually doing? It runs R underneath and it is going to give you the estimated power of the t-test under both conditions. The first one is calculated under the assumption that your data are bivariate normally distributed. This is just a direct use of the pwr.r.test function of the pwr R package. Should give you answers very close or exactly the same as G*Power. I chose to use it for sake of comparison.

What comes out is something that looks like this:

So, on top, you’re going to get the power as if you were using G*Power (or the pwr R package) which is exact and needs no simulation because we have closed-form expressions for it. On the bottom you are going to get the approximated power using simulations. Remember, when working with non-normal data you can’t always expect power to be lower, as in this case. Sometimes it may be higher and sometimes it won’t change much. Yes, at some point (if your sample size is large enough) the distribution of your data will matter very little. But, in the meantime, at least you can use this one to guide you!

Finally, here’s the link for the shiny web app:

# Sylvester’s criterion to diagnose Heywood cases.

So… here’s a trick I use sometimes that I’ve never seen anywhere else and it may be helpful to some people.

If you work with latent variable models, Factor Analysis, Structural Equation Modelling, Item Response Theory, etc. there’s a good chance that you have either encountered or have seen some version of a warning about a covariance matrix being “non positive definite”. This is an important warning because the software is telling you that your covariance matrix is not a valid covariance matrix and, therefore, your analysis is suspect. Usually, within the world of latent variables we call these types of warnings Heywood cases .

Now, most Heywood cases are very easy to spot because they pertain to one of two broad classes: negative variances or correlations greater than 1. When you inspect your matrix models and see either of those two cases, you know exactly which variable is giving you trouble. Thing is (as I found out a few years ago), there are other types of Heywood cases that are a lot more difficult to diagnose. Consider the following matrix that I once got helping a student with his analysis:

 ............space    lstnng    actvts    prntst    persnl    intrct    prgrmm space       1.000 lstnng      0.599    1.000 actvts      0.706    0.646     1.000 prntst      0.702    0.459     0.653      1.000 persnl      0.591    0.582     0.844      0.776     1.00 intrct      0.627    0.964     0.501      0.325     0.639    1.000 prgrmm      0.493    0.602     0.981      0.687     0.944    0.642      1.000 

This is the model-implied correlation matrix I obtained through the analysis which gave a Heywood case warning. The student was a bit puzzled because, although we had reviewed this type of situations in class, we had only mentioned the case of negative variances or correlations greater than one. Here he neither had a negative variance nor a correlation greater than one… but he still got a warning for non-positive definiteness.

My first reaction was, obviously, to check the eigenvalues of the matrix and, lo and behold, there it was: [1] 5.01377877 1.00744933 0.62602056 0.30393170 0.16671742 0.01317704 -0.13107483

So… yeah. This was, indeed, an invalid correlation/covariance matrix and we needed to further diagnose where the problem was coming from… but how? Enter our good friend, linear algebra.

If you have ever taken a class in linear algebra beyond what’s required in a traditional methodology/statistics course sequence for social sciences, you may have encountered something called the minor of a matrix. Minors are important because they’re used to calculate the determinant of a matrix. They’re also important for cases like this one because they break down the structure of a matrix into simpler components that can be analyzed. Way in the back of my mind I remembered from my undergraduate years that positive-definite matrices had something special about their minors. So when I came home I went through my old, OLD notes and found this beautiful theorem known as Sylvester’s criterion:

A Hermitian matrix is positive-definite if and only if  all of the leading principal minors have positive determinant.

All covariance matrices are Hermitian (the subject for another blogpost) so we’re only left to wonder what is a principal minor. Well, if you imagine starting at the [1,1] coordinate of a matrix (so really the upper-left entry) and going downwards diagonally expanding one row and column at a time you’d end up with the principal minors. A picture makes a lot more sense for this:

So… yeah. The red square (so the q11 entry) is the first principal minor. The blue square (a 2 x 2 matrix) is the 2nd principal minor, the yellow square (a 3 x 3 matrix) is the 3rd principal minor and on it goes until you get the full n x n matrix. For the matrix Q to be positive-definite, all the n-1 principal minors need to have positive determinant. So if you want to “diagnose” your matrix for positive definiteness, all you need to do is start from the upper-left corner and check the determinants consecutively until you found one that is less than 0. Let’s start from the previous example. Notice that I’m calling the matrix ‘Q’: 1 > det(Q[1:2,1:2]) [1] 0.641199 > det(Q[1:3,1:3]) [1] 0.2933427 > det(Q[1:4,1:4]) [1] 0.01973229 > det(Q[1:5,1:5]) [1] 0.003930676 > det(Q[1:6,1:6]) [1] -0.003353769 

The first [1,1] corner is a given (it’s a correlation matrix so it’s 1 and it’s positive). Then we move downwards the 2×2 matrix, the 3×3 matrix… all the way to the 5×5 matrix. By then the determinant of Q is very small so I suspected that whatever the issue might be, it had to do with the relationship of the variables  “persnl”,  “intrct” or “prgrmm”. The final determinant pointed towards the culprit. Whatever problem this matrix exhibited, it had to do with the relationship between “intrct” and”prgrmm”.

Once I pointed out to the student that I suspected the problem was coming from either one of these two variables, a careful examination revealed the cause: The “intrct” item was a reverse-coded item but, for some reason, several of the participants respondents were not reverse-coded. So you had a good chunk of the responses to this item pointing to one direction and a smaller (albeit still large) number pointing to the other direction. The moment this item was full reverse-coded the non-positive definite issue disappeared.

I guess there are two lessons to this story: (1) Rely on the natural structure of things to diagnose problems and (2) Learn lots of linear algebra 🙂

# The Spearman correlation doesn’t need to tell you anything about the Pearson correlation

I… am a weird guy (but then again, that isn’t news, hehe). And weird people sometimes have weird pet-peeves. Melted cheese on Italian food and hamburgers? Great! I’ll take two. Melted cheese anywhere else? Blasphemy! That obviously extends to Statistics. And one of my weird pet peeves (which was actually strong enough  to prompt me to write it as a chapter for my dissertation and a subsequent published article) is when people conflate the Pearson and the Spearman correlation.

The theory of what I am going to talk about is developed in said article but when I was cleaning more of my computer files I found an interesting example that didn’t make it there. Here’s the gist of it:

I really don’t get why people say that the Spearman correlation is the ‘robust’ version or alternative to the Pearson correlation. Heck, even if you simply google the words Spearman correlation the second top hit reads

Spearman’s rank-order correlation is the nonparametric version of the Pearson product-moment correlation

When I read that, my mathematical mind immediately goes to “if this is the non-parametric version of the Pearson correlation, that means it also estimates the same population parameter”. And honest to G-d (don’t quote me on that one, though. But I just *know* it’s true) I feel like the VAST MAJORITY of people think exactly that about the Spearman correlation. And I wouldn’t blame them either… you can’t open an intro textbook for social scientists that doesn’t have some dubious version of the previous statement. “Well” – the reader might think- “if that is not true then why aren’t more people saying it?” The answer is that, for better or worse, this is one of those questions that’s very simple but the answer is mathematically complicated. But here’s the gist of it (again, for those who like theory like I do, read the article).

The Spearman rank correlation is defined, in the population, like this:

where the $u_i$ are uniformly-distributed random variables and the $C(\cdot)$ is the copula function that relates them (more on copulas here.) By defining the Spearman rank correlation in terms of the lower-dimensional marginals it co-relates (i.e. the u’s) and the copula function, it becomes apparent that the overlap with the Pearson correlation depends entirely on what $C(u_x,u_y)$ is. Actually, it is not hard to show that if $C(\cdot)$ is a Gaussian copula, and the marginals are normal, then the following identity can be derived:

$\rho_S=\frac{6}{\pi}\sin^{-1}\left ( \frac{\rho}{2} \right )$

That relates the Pearson correlation $\rho$ to the Spearman correlation. We’ve known this since the times of Pearson because he came up with it (albeit not explicitly) and called it the “grade correlation”. But from this follows the obvious. An identity such as the one described above need not exist. There could be copula functions for which the Spearman and Pearson correlation have a crazy, wacky relationship… and that is what I am going to show you today.

I don’t remember why but I didn’t include this example in the article, but it is a very efficient one. It shows a case where the Spearman correlation is close to 1 but the Pearson correlation is close to 0 (obviously within sampling error).

Define $Z \sim N(0, 0.1)$ , $X=Z^{201}$ and $Y= e^{Z}$.  if you use R to simulate a very large sample (say 1 million) then you can find the following Spearman and Pearson correlations:

 N <- 1000000 Z <- rnorm(N, mean=0, sd=0.1) X <- Z^201 Y <- exp(Z)  > cor(X,Y, method="pearson") [1] 0.004009381  > cor(X,Y, method="spearman") [1] 0.9963492 

The trick for this example is to notice that the rate of change of  $X$ is microscopic and oscillating around 0 whereas the change in $Y$ is fairly small and oscillating around 1.

In any case, the point being that even without being necessarily too formal, it’s not overly difficult to see that the Spearman correlation is its own statistic and estimates its own population parameter that may or may not have anything to do with the Pearson correlation, depending on the copula function describing the bivariate distribution.

# Power Analysis for Multilevel Logistic Regression

## ::UPDATE::

#### The relationship between statistical power and predictor distribution in multilevel logistic regression: a simulation-based approach

WARNING (1): This app can take a little while to run. Do not close your web browser unless it gives you an error. If it appears ‘stuck’ but you haven’t got an error it means the simulation is still running on the background.

WARNING (2): If you keep getting a ‘disconnected  from server’ error, close down your browser and open a new window. If the problem still persists that means too many people have tried to access it during the day and the server has shut down. This app is hosted on a free server and it can only accommodate a certain number of people every day.

This app will perform computer simulations to estimate power for multilevel logistic regression models allowing for continuous or categorical covariates/predictors and their interaction. The continuous predictors come in two types: normally distributed or skewed (i.e. χ2  with 1 degree of freedom). It currently only supports binary categorical covariates/predictors (i.e. Bernoulli-distributed) but with the option to manipulate the probability parameter p to simulate imbalance of the groups.

The app will give you the power for each individual covariate/predictor  AND the variance component for the intercept (if you choose to fit a random-intercept model) or the slope (if you choose to fit a model with both a random intercept and a random slope). It uses the Wald test statistic for the fixed effect predictors and a 1-degree-of-freedom likelihood-ratio test for the random effects (← yes, I know this is conservative but it’s the fastest one to implement).

When you open the app, here’s how it looks:

What **you**, as the user, need to provide is the following:

The Level 1 and Level 2 sample sizes. If I were to use the ubiquitous example of “children in schools” the Level 1 sample would be the children (individuals within a cluster) and the Level 2 sample would be the schools  (number of clusters). For demonstration purposes here I’m asking for groups of 50 ‘children’ in 10 ‘schools’ for a total sample size of 50×10 = 500 children.

The variance for the random effects. You can either choose to fit an intercept-only model (so no variance of the slope) or a random intercept AND random slope model. You cannot fit a random-slope only model here and you cannot set the variances at 0 to fit a single-level logistic regression (there’s other software to do power analysis for single-level logistic regression). At least the variance of the intercept needs to be specified. Notice that the app defaults to an intercept-only model and under ‘Select Covariate’ it will say ‘None’. That changes when you click on the drop-down menu where it gives you the option of which random slope do you want. Notice that you can only choose one predictor to have a random slope. Will work on the general case in the future.

The number of covariates (or predictors) which I believe is pretty self-explanatory. Just notice that the more covariates you add, the longer it will take for the simulation to run. The default in the app is 2 covariates.

This would be the core of the simulation engine because the user needs to specify:

• Regression coefficients (‘Beta’).  This space lets the user specify the effect size for the regression coefficients under investigation. The default is 0.5 but that can be changed to any number. In the absence of any outside guidance, Cohen’s small-medium-large effect sizes are recommended. Remember that the regression coefficient for binary predictors is conceptualized as a standardized mean difference so it should be in Cohen’s d metric.
• Level of the predictor (‘Level’). It only supports 2-level models so the options are ‘1’ or ‘2’. This section indicates whether a predictor belongs to the Level 1 sample (e.g. the ‘children’) or the Level 2 sample (e.g. the ‘school). Notice that whichever predictor gets assigned a random slope MUST also be selected as Level 1. Otherwise the power analysis results will not make sense. It currently only supports one predictor at the Level 1 with a random slope. Other predictors can be included at Level 1 but they won’t have the option for a random slope component.
• Distribution of the covariates (‘Distribution’).  Offers 3 options: normally-distributed, skewed  (i.e. χ2  with 1 degree of freedom or a skew of about √8) and binary/Bernoulli-distributed. For the binary predictor the user can change the population parameter and create imbalance between the groups. So, for instance, if p=0.3 then 30% of the sample would belong to the group labelled as ‘1’ and 70% to the group labelled as ‘0’. The default for this option is 0.5 to create an even 50/50 split.
• Intercept (‘Intercept Beta’).  Lets the user define the intercept for the regression model. The default is 0 and I wouldn’t recommend changing it unless you’re making inferences about the intercept of the regression model.

Once the number of covariates has been selected, the app will offer the user all possible 2-way interaction effects irrespective of the level of the predictor and distribution characteristics. The user can select whichever 2-way interaction is of interest and assign an effect size/regression coefficient (i.e. ‘Beta’). The app will use this effect size to calculate power. Notice that the distribution of the interaction is fully defined by the distribution of its constituting main effects.

The number of datasets generated using the population parameters previously defined by the researcher. The default is 10 but I would personally recommend a minimum of 100. The larger the number of replications the more accurate the results will be but also the longer the simulation will take.

The simulated power is calculated as the proportion of statistically significant results out of the number of simulated datasets and will be printed here. Notice the time progress bar indicating that the simulation is still running. For a 2-covariate model with both a random effect for the intercept and the slope the simulation took almost 3 min to run. Expect longer waiting times if the model has lots of covariates.

This is what a sample of a full power analysis looks like. The estimated power can be found under the column ‘Power’. The column labelled ‘NA’ shows the proportion of models that did not converge. In this case, all models converged (there are 0s all throughout the NA column) but the power of the fixed and random effects is relatively low with the exception of the power for the variance of the random intercept. In this example one would need to either increase the effect size from 0.5 to something larger or increase the Level 1 and Level 2 sample sizes in order to obtain acceptable power levels of 80%. You can either download your power analysis results as a .csv file or copy-paste them by clicking on the appropriate button.

Finally, here is the link for the shiny web app:

# Ordinal Alpha and Parallel Analysis

This shiny app will:

– Give you the polychoric (or tetrachoric, in case of binary data) correlation matrix

– Do Parallel Analysis and a scree plot based on the polychoric (or tetrachoric) correlation matrix

– Calculate ordinal alpha as recommended in:

Zumbo, B. D., Gadermann, A. M., & Zeisser, C.. (2007). Ordinal Versions of Coefficients Alpha and Theta For Likert Rating Scales. Journal of Modern Applied Statistical Methods, 6, 21-29.

It currently takes in certain SPSS files (so .sav file extensions from older versions of SPSS, say around 2013 or less), Microsoft Excel files (so .xls file extensions) and comma-delimited files (so .csv extensions). If your data is in none of those files, please change it before using the app (it’s super easy). or it will give you an error. Also notice that the app will use ALL of the variables in the file uploaded, so make sure you upload a file that only has the variables (test items in most cases) which you want to correlate/calculate alpha for. You’ll need to provide a clean dataset for it to work. So if you have missing values, you’ll need to manually remove them before submitting it. If there are outliers, those need to be dealt with before using the app.

Please notice that in accordance to research, if you have 8 (or more) Likert responses, the app will give you an error saying you have enough categories to safely treat your variables as continuous, so you don’t really need to use this app. You can see why in Rhemtulla, Brosseau-Liard & Savalei (2012).