I am used to knowing "degrees of freedom" as $n - r$, where you have the linear model $$\mathbf{y} = \mathbf{X}\boldsymbol{\beta} + \boldsymbol{\epsilon}$$ with $\mathbf{y} \in \mathbb{R}^n$, $\mathbf{X} \in M_{n \times p}(\mathbb{R})$ the design matrix with rank $r$, $\boldsymbol{\beta} \in \mathbb{R}^p$, $\boldsymbol{\epsilon} \in \mathbb{R}^n$ with $\boldsymbol{\epsilon} \sim \mathcal{N}(\mathbf{0}, \sigma^2 \mathbf{I}_n)$, $\sigma^2 > 0$.

From what I recall of elementary statistics (i.e., pre-linear models with linear algebra), the degrees of freedom for the matched-pairs $t$-test is the number of differences minus $1$. So this would entail $\mathbf{X}$ having rank 1, perhaps. Is this correct? If not, why is $n-1$ the degrees of freedom for the matched-pairs $t$-test?

To understand the context, suppose I have a mixed-effects model $$y_{ijk} = \mu_i + \text{ some random effects} + e_{ijk}$$ where $i = 1, 2$, $j = 1, \dots, 8$, and $k = 1, 2$. There is nothing special about $\mu_i$ other than that it's a fixed effect, and $e_{ijk} \overset{iid}{\sim}\mathcal{N}(0, \sigma^2_e)$. I'm assuming that the random effects are irrelevant to this problem, since we only care about the fixed effects in this case.

I would like to provide a confidence interval for $\mu_1 - \mu_2$.

I have already shown that $\bar{d}_\cdot = \dfrac{1}{8}\sum d_j$ is an unbiased estimator of $\mu_1 - \mu_2$, where $d_j = \bar{y}_{1j\cdot} - \bar{y}_{2j\cdot}$, $\bar{y}_{1j\cdot} = \dfrac{1}{2}\sum_{k}y_{1jk}$, and $\bar{y}_{21\cdot}$ is defined similarly. The point estimate $\bar{d}_{\cdot}$ has been computed.

I have already shown that $$s^2_d = \dfrac{\sum_{j}(d_j - \bar{d}_{\cdot})^2}{8-1}$$ is an unbiased estimator of the variance of $d_j$, and thus, $$\sqrt{\dfrac{s^2_d}{8}}$$ is the standard error of $\bar{d}_{\cdot}$. This has been computed.

Now the last part is figuring out the degrees of freedom. For this step, I usually try to find the design matrix - which obviously has rank 2 - but I have the solution to this problem, and it says that the degrees of freedom is $8-1$.

In the context of finding the rank of a design matrix, why are the degrees of freedom $8-1$?

Edited to add: Perhaps helpful in this discussion is how the test statistic is defined. Suppose I have a parameter vector $\boldsymbol{\beta}$. In this case, $$\boldsymbol{\beta} = \begin{bmatrix} \mu_1 \\ \mu_2 \end{bmatrix}$$ (unless I'm missing something entirely). We are essentially performing the hypothesis test $$\mathbf{c}^{\prime}\boldsymbol{\beta} = 0$$ where $\mathbf{c}^{\prime} = \begin{bmatrix} 1 & -1 \end{bmatrix}$. Then, the test statistic is given by $$t = \dfrac{c^{\prime}\hat{\boldsymbol{\beta}}}{\sqrt{\hat{\sigma}^2c^{\prime}(\mathbf{X}^{\prime}\mathbf{X})^{-1}\mathbf{c}}}$$ which would be tested against a central $t$-distribution with $n - r$ degrees of freedom, where $\mathbf{X}$ is the design matrix as above, and $$\hat{\sigma}^2 = \dfrac{\mathbf{y}^{\prime}(\mathbf{I}-\mathbf{P}_{\mathbf{X}})\mathbf{y}}{n-r}$$ where $\mathbf{P}_{\mathbf{X}} = \mathbf{X}(\mathbf{X}^{\prime}\mathbf{X})^{-1}\mathbf{X}^{\prime}$.


2 Answers 2


The matched-pairs $t$-test with $n$ pairs is actually just a one-sample $t$-test with a sample of size $n$. You have $n$ differences $d_1,\ldots,d_n$, and these are i.i.d. and normally distributed. $$ \begin{array}{ccccc} \begin{bmatrix} d_1 \\ \vdots \\ d_n \end{bmatrix} & = & \begin{bmatrix} \bar d \\ \vdots \\ \bar d \end{bmatrix} & + & \begin{bmatrix} d_1 - \bar d \\ \vdots \\ d_1 - \bar d \end{bmatrix} \\[10pt] n \text{ d.f.} & & 1 \text{ d.f.} & & (n-1) \text{ d.f.} \end{array} $$ The first column after $\text{“}{=}\text{''}$ has $1$ degree of freedom because of the linear constraint that says all entries are equal; the second has $n-1$ degrees of freedom because of the linear constraint that says the sum of the entries is $0$.

  • $\begingroup$ So, in other words, the reason why we have $n-1$ degrees of freedom here has nothing to do with the linear model $\mathbf{y}=\mathbf{X}\boldsymbol{\beta}+\boldsymbol{\epsilon}$? $\endgroup$ May 1, 2016 at 14:15
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    $\begingroup$ It does have to do with that model, where the matrix $\mathbf X$ is a column of $1$s and $\boldsymbol{\beta}$ is a $1\times1$ matrix whose only entry is the difference between the two population means. $\qquad$ $\endgroup$ May 1, 2016 at 14:19
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    $\begingroup$ Aha! So your $\mathbf{y}$ vector would be that vector of $d_i$s, correct? Thank you very much! I can't believe how hard it has been to find an answer on this! $\endgroup$ May 1, 2016 at 14:20
  • $\begingroup$ Yes. It's the vector of observed differences in the $n$ matched pairs. $\qquad$ $\endgroup$ May 1, 2016 at 14:21

Many, many thanks to Michael Hardy for answering my question.

The idea is this: let $$\mathbf{y} = \begin{bmatrix} d_1 \\ \vdots \\ d_n \end{bmatrix}$$ and $\boldsymbol{\beta} = [\mu_1 - \mu_2]$. Then our linear model is then $$\mathbf{y} = \mathbf{1}_{n \times 1}\boldsymbol{\beta} + \boldsymbol{\epsilon}$$ where $\mathbf{1}_{n \times 1}$ is the $n$-vector of all ones, and $$\boldsymbol{\epsilon} = \begin{bmatrix} \epsilon_1 \\ \vdots \\ \epsilon_n \end{bmatrix} \sim \mathcal{N}(\mathbf{0}, \sigma^2\mathbf{I}_n)\text{.}$$ Obviously $\mathbf{X} = \mathbf{1}_{n \times 1}$ has rank $1$, so then we have $n-1$ degrees of freedom.

How do we know to set $\boldsymbol{\beta}$ equal to $[\mu_1 - \mu_2]$? Recall that $$\mathbb{E}[\mathbf{y}] = \mathbf{X}\boldsymbol{\beta}$$ and as it can be easily seen, $\mathbb{E}[d_j] = \mu_1 - \mu_2$ for all $j$. Given our $\mathbf{X}$, it is obvious what $\boldsymbol{\beta}$ should be. This is because $$\mathbb{E}[\mathbf{y}] = \mathbb{E}\left[\begin{bmatrix} d_1 \\ \vdots \\ d_n \end{bmatrix} \right] = \begin{bmatrix} \mathbb{E}[d_1] \\ \vdots \\ \mathbb{E}[d_n] \end{bmatrix} = \begin{bmatrix} \mu_1 - \mu_2 \\ \vdots \\ \mu_1 - \mu_2 \end{bmatrix} = \mathbf{X}\boldsymbol\beta = \mathbf{1}_{n \times 1}\boldsymbol\beta = \begin{bmatrix} 1 \\ \vdots \\ 1 \end{bmatrix}\boldsymbol\beta$$ so $\boldsymbol\beta$ should be a $1 \times 1$ matrix with $\boldsymbol\beta = [\mu_1 - \mu_2]$.

Set $\mathbf{c}^{\prime} = [1]$. Then our hypothesis test is $$H_0: \mathbf{c}^{\prime}\boldsymbol{\beta} = 0\text{.}$$ Our test statistic is thus $$\dfrac{\mathbf{c}^{\prime}\hat{\boldsymbol{\beta}}}{\sqrt{\hat{\sigma}^2\mathbf{c}^{\prime}\left(\mathbf{X}^{\prime}\mathbf{X}\right)^{-1}\mathbf{c}}}\text{.}$$ We have $$\hat{\sigma}^2 = \dfrac{\mathbf{y}^{\prime}(\mathbf{I}-\mathbf{P}_{\mathbf{X}})\mathbf{y}}{n-r(\mathbf{X})}\text{.}$$ After some work, it can be shown that $$\mathbf{P}_\mathbf{X} = \mathbf{P}_{\mathbf{1}_{n \times 1}} = \mathbf{1}_{n \times 1}\left(\dfrac{1}{n}\right)\mathbf{1}^{\prime}\text{.}$$ It can also be shown that $\mathbf{I}-\mathbf{P}_{\mathbf{X}}$ is symmetric and idempotent. So, $$\begin{align} \hat{\sigma}^2 &= \dfrac{\mathbf{y}^{\prime}(\mathbf{I}-\mathbf{P}_{\mathbf{X}})\mathbf{y}}{n-r(\mathbf{X})} \\ &= \dfrac{\mathbf{y}^{\prime}(\mathbf{I}-\mathbf{P}_{\mathbf{X}})^{\prime}(\mathbf{I}-\mathbf{P}_{\mathbf{X}})\mathbf{y}}{n-r(\mathbf{X})} \\ &= \dfrac{\|(\mathbf{I}-\mathbf{P}_{\mathbf{X}})\mathbf{y}\|^{2}}{n-r(\mathbf{X})} \\ &=\dfrac{\left\|\left[\mathbf{I}-\mathbf{1}_{n \times 1}\left(\dfrac{1}{n}\right)\mathbf{1}^{\prime}\right]\mathbf{y} \right\|^2}{n-1} \\ &= \dfrac{\left\|\begin{bmatrix} d_1 \\ \vdots \\ d_n \end{bmatrix} - \begin{bmatrix} \bar{d}_\cdot \\ \vdots \\ \bar{d}_\cdot \end{bmatrix} \right\|^2}{n-1} \\ &= \dfrac{\sum_{i=1}^{n}(d_i-\bar{d}_{\cdot})^2}{n-1} \\ &= s^2_d \end{align}$$ and $$\mathbf{X}^{\prime}\mathbf{X} = \mathbf{1}_{n \times 1}^{\prime}\mathbf{1}_{n \times 1} = n$$ which obviously has inverse $1/n$, thus giving a test statistic $$\dfrac{\hat\mu_1-\hat\mu_2}{\sqrt{s^2_d/n}}$$ which would be tested on a central $t$-distribution with $n - 1$ degrees of freedom as desired.


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