# Question about only centering a matrix once in the covariance matrix

From probability, we have $$\Sigma_x = \frac{\left(HX\right)^T\left(HX\right)}{N}$$ Where $$\Sigma_X$$ is the covariance matrix for a data matrix $$X$$. $$X$$ is constructed by putting all the datapoints in rows, so the first row is the first datapoint, the second row is the second collected data point and so forth. $$H$$ is the centering matrix $$I - \frac{11^T}{N}$$

Simplifying $$\Sigma_X$$, $$\frac{\left(HX\right)^T\left(HX\right)}{N} = \frac{X^TH^T H X }{N} = \frac{X^T H ^2 X}{N} = \frac{X^THX}{N}$$

This both makes sense and doesn't make sense. At first glance, centering a matrix once and then doing so again does nothing the second time around. Matrix $$H$$ is idempotent. As a result when we lose the squared above the H, I'm okay with that.

However digging a little deeper, it doesn't make sense that only one of my data matrices is centered when I should have two centered data matrices to properly compute covariance.

Going to the 1D case, $$E[(X - \mu_{x})^2] \neq E[X(X - \mu_{x})]$$

Here it doesn't make sense for only one of my $$X$$ random variables to be centered while the other one isn't.

• Your last inequality is not true. Indeed, this checks out: \begin{align} \mathbb{E}\big[(X - \mu_{x})^2\big] &= \mathbb{E}\big[X^2 - 2X\mu_{x}+\mu_x^2\big]\\ &= \mathbb{E}\big[X^2\big] - 2\mathbb{E}\big[X\big]\mu_{x}+\mu_x^2\\ &= \mathbb{E}\big[X^2\big] - 2\mu_{x}^2+\mu_x^2\\ &= \mathbb{E}\big[X^2\big] - \mu_x^2\\ &= \mathbb{E}\big[X^2\big]-\mathbb{E}\big[X\big]\mu_x\\ &= \mathbb{E}\big[X^2-X\mu_x\big]\\ &= \mathbb{E}\big[X(X - \mu_{x})\big] \end{align} Nov 8, 2023 at 21:48
• @cjferes Can you explain why that true intuitively? It's very odd to say that $(x-1)*(x - \mu_x)$ is the same as, say, $(x + 999999)*(x - \mu_x)$. We know that $E[(X - \mu_x)^2]$ is the the "average" deviation of random variable X from it's mean. We square the factor inside the expectation because we don't want positive deviations to be canceled out by negative deviations from the mean. Is there a similar way I can understand the $E[X(X - \mu_x)]$ formulation? Nov 9, 2023 at 19:38
• @BigBear $(x-1) (x-\mu_x)$ and $(x+99999)(x-\mu_x)$ are not the same but their means/averages are ! Nov 11, 2023 at 16:21
• Also, be careful with the statement about squaring. If we only care about the sign of the difference, we would use absolute value. Squaring also gives more weight to large deviations and smaller weight to smaller deviations, which changes how you measure dispersion. Nov 11, 2023 at 17:23
• Can you evaluate what is the question you have encountered? Since reading the comments after reading the posts, I'm not sure where is the question... Nov 12, 2023 at 19:42

Consider the 1D case. I suspect you are implicitly thinking as follows: the function $$g(s,t)=E(X-s)(X-t)$$ has gradient $$\nabla g(s,t) = (t-EX, s-EX)$$, so it achieves a unique minimum at $$(s,t)=(EX,EX)$$. The minimum value is the variance $$\mathrm{Var}(X)=g(EX,EX)$$.

However, this condition is extremely strong: it means that $$g(s,t)\ge g(EX,EX)$$ for all $$(s,t)\in\mathbb R^2$$. Instead, let's first restrict $$g$$ to a line passing through $$(EX,EX)$$, by considering the function $$f(s)=g(s,EX)$$. This changes the situation significantly: we have $$f'(s)=0$$ for all $$s$$, so $$f$$ is constant. (Of course, we know that $$f\equiv\mathrm{Var}(X)$$, but the main task should be intuitively understanding why $$f$$ is constant.)

Let $$Y=X-\mathbb{E}\{X\}=X-\mu_X$$ be the centered version of $$X$$. Your question is that you don't "see" why is the case that $$\mathrm{Var}\{Y\}=\mathrm{Cov}\{X,Y\}=\mathrm{Var}\{X\}$$

You have realized by now that, with $$H$$ being idempotent, this is actually the case. Noting that $$\mathbb{E}\{Y\}=\mu_Y=0$$, we also can see that \begin{align} \mathrm{Var}\{Y\}&=\mathbb{E}\{(Y-\mu_Y)^2\}=\mathbb{E}\{Y^2\}=\mathbb{E}\{(X-\mu_X)^2\}=\mathrm{Var}\{X\} \end{align} and furthermore, \begin{align} \mathbb{E}\big[(X - \mu_{x})^2\big] &= \mathbb{E}\big[X^2 - 2X\mu_{x}+\mu_x^2\big]\\ &= \mathbb{E}\big[X^2\big] - 2\mathbb{E}\big[X\big]\mu_{x}+\mu_x^2\\ &= \mathbb{E}\big[X^2\big] - 2\mu_{x}^2+\mu_x^2\\ &= \mathbb{E}\big[X^2\big] - \mu_x^2\\ &= \mathbb{E}\big[X^2\big]-\mathbb{E}\big[X\big]\mu_x\\ &= \mathbb{E}\big[X^2-X\mu_x\big]\\ &= \mathbb{E}\big[X(X - \mu_{x})\big]\\ &= \mathbb{E}\big[XY\big]\\ &= \mathbb{E}\big[XY\big]-\mu_X\mu_Y\\ &=\mathrm{Cov}\{X,Y\} \end{align}

This can also be understood by the fact that $$\mathrm{Cov}\{U,V\}=\mathrm{Cov}\{U+a,V+b\}$$, with $$a,b$$ constants, for all random variables $$U,V$$. If we use this property with $$U=X$$, $$a=0$$, $$V=X$$ and $$b=\mu_X$$, we obtain again the same result.

Why am I relating the result to a covariance? To attempt an intuitive explanation beyond these analytical proofs.

Recall that the covariance of $$U$$ and $$V$$ is a measure of the joint variability of these two variables, i.e., how they change with respect to one another. Shifting one or both them by a constant term does not change the behavior of this (relative) joint variability.

Now, $$\mathrm{Var}\{U\}=\mathrm{Cov}\{U,U\}$$, i.e. variance is a measure of variability of a random variable against itself, a dispersion with respect to its mean (in quadratic terms), but is also a covariance. Hence, it does not change by shifting one or both copies of $$U$$ in $$\mathrm{Cov}\{U,U\}$$ by any constant, in particular its mean.