09/09/2020

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The following topics will be covered in this lecture:

- Gaussian error regression models
- Estimation by Maximum Likelihood
- Gaussian correlation models

- Recall now our standard regression equation, \[ \begin{align} Y_i = \beta_0 + \beta_1 X_i + \epsilon_i. \end{align} \]
- Up to now we have only made assumptions about the first two moments of the variation in the signal, \( \epsilon_i \), and the decorrelation of cases:
- \( \mathbb{E}[ \epsilon_i ] = 0 \) for each \( i \);
- \( \mathbb{E}[ \epsilon_i^2 ] = \sigma^2 \) for each \( i \); and
- \( \mathbb{E}[ \epsilon_i \epsilon_j ] = 0 \) for every \( i\neq j \).
- Regardless of the actual distribution of the error/ variation \( \epsilon_i \), the Gauss-Markov theorem holds and the solution by least squares is the BLUE.
- However, to create confidence intervals and perform uncertainty quantification, we will need to assume some additional structure.
- It is often assumed that the error distribution for \( \epsilon_i \) is Gaussian for several reasons:
- Gaussian errors simplify the regression analysis significantly;
- Gaussian distributions are quite common in real applications;
- it can be shown that linear combinations of Gaussian random variables are also Gaussian so that, in a linear model, many other quantities will also be Gaussian; and finally
- even when errors are not strictly Gaussian, a Gaussian distribution is often a “good” approximation.
- The notion of a Gaussian distribution being a “good” approximation is formalized with the Central Limit Theorem, which we now recall.

- Suppose we have a sequence of random variables, \( \left\{x_i \right\}_{i=1}^\infty \) which are independent, identically distributed
**from any distribution**, such that for all \( i \)- \( \mathbb{E}[x_i] = \mu \); and
- \( \mathrm{var}(x_i) = \sigma^2 \), for some finite \( \sigma \).

- For each \( n\geq 1 \) define the sample-based mean of the sequence \( \left\{x_i \right\}_{i=1}^n \) \[ \begin{align} \overline{x}_n \triangleq \frac{1}{n} \sum_{i=1}^n x_i \end{align} \]
- Then, as the number of random variables \( n\rightarrow \infty \), the sample-based means \[ \begin{align} \sqrt{n}\left(\overline{x}_n - \mu\right) {\xrightarrow {d}} N(0, \sigma^2) \end{align} \] where the convergence is in distribution.
- Put another way, for \( n \) sufficiently large, \( \overline{x}_n \) has
**approximately**a \( N\left(\mu, \frac{\sigma^2}{n}\right) \) distribution. - Heuristically,
**for large sample sizes**, we can produce confidence intervals for the mean of the unbiased estimator (e.g., the true \( \beta_0,\beta_1 \)),- from the sample based estimate (the least squares solution \( \hat{\beta}_0 ,\hat{\beta}_1 \)) with a Gaussian assumption;
- this is a
**good approximation**if not strictly accurate.

- The “goodness” of this approximation can be made more formal in another course.

- The Gaussian assumption will be introduced on top of our existing assumptions as follows:
- We suppose that we have a signal in the data, linear in the parameters \( \beta_0,\beta_1 \), \[ Y_i = \beta_0 + \beta_1 X_i + \epsilon_i \]
- \( X_i \) is a known constant, while \( \beta_0,\beta_1 \) are unkown but fixed values for all \( i \).
- \( \epsilon_i \) are drawn independently and identically distributed (iid) as, \[ \epsilon_i \sim N\left(0, \sigma^2\right) \] for each \( i \).
- Independence of the \( \epsilon_i \) implies that the cases are once again uncorrelated.
- Additionally, as we have assumed the form of the distribution as Gaussian, and the form of the first two moments, we have entirely parametrized the distribution of the variation.

- As was mentioned before, Gaussianity thus simplifies the regression analysis significantly.
- It is easy to construct confidence intervals for the mean of a sample-based estimate of the mean of a Gaussian random variable from the Student t distribution.
- Likewise, the t-test derived from the Student t distribution can be used to perform hypothesis tests.
- We will return to both of these points when we introduce multiple regression more formally.
- For the moment, we will return to the addendum of the Gauss-Markov theorem:
- If in addition, \( \epsilon\sim N\left(0,\sigma^2\right) \), then the solution by least squares is the “maximum likelihood estimator”.

- A concept widely used in statistical inference is a “likelihood” function.
- Consider a probability density, denoted \( p_\theta(y) \), which depends on the value of the parameter \( \theta \).
- Suppose we are investigating some random variable \( Y \), for which there is some “true” parameter \( \theta_0 \), such that \[ P(a \leq Y \leq b)= \int_a^b y p_{\theta_0}(y) \mathrm{d}y. \]
- We assume that even though \( \theta_0 \) is not known,
**we can evaluate the density**\[ \begin{align} p_\theta(Y=y) \end{align} \] for any choice of \( \theta \) and some observed piece of data \( y \). - Then, the likelihood of \( \theta \) based on an observed \( y \) is defined
\[ \begin{align}
\mathcal{L}(\theta \vert y) = p_\theta(Y=y)
\end{align} \]
where we evaluate the probability of \( Y \) attaining an observed piece of data \( y \) given a
**choice**of \( \theta \). - The “likelihood” is thus a measure of how well does our choice of parameter make the distribution describe the observed data.
- Maximum likelihood estimation is thus the process of finding a \( \hat{\theta} \) which maximizes the likelihood,
- i.e., \( \hat{\theta} \) which maximizes the probability density for the observed data \( y \) given the density \( p_\theta \).

Courtesy of: Kutner, M. et al. Applied Linear Statistical Models 5th Edition

- Suppose that we have three samples \( Y_1,Y_2,Y_3 \) from what we know is a population that is Gaussian distributed.
- Moreover, we suppose that while we know the standard deviation \( \sigma=10 \), we do not know the true mean \( \mu \).
- Let \( \mathrm{exp}\left\{x\right\} \) denote the function \( e^x \).
- By varying the choice of \( \mu \), we can evaluate the density of any particular sample with the Gaussian density
\[ f_j = \frac{1}{\sqrt{2\pi}10}\mathrm{exp}\left\{-\frac{1}{2}\left(\frac{Y_j - \mu}{10}\right)^2\right\}, \]
denoted \( f_j \) for sample \( j \), based on the
**choice**of \( \mu \). - We suppose that the observed data points are \( Y_1 = 250, Y_2=265,Y_3=259 \)
**Q:**given the above data points, and the associated graph to the left, does \( \mu=230 \) appear to be the “most likely” choice for the true mean?

Courtesy of: Kutner, M. et al. Applied Linear Statistical Models 5th Edition

- Intuitively, we can tell that there are better choices for the “center of mass” of the data, given by our choice of \( \mu \).
- One particular choice may be to set \( \mu= 259= Y_3 \) as on the left.
- Indeed, we can compare the values for the density function for each choice of \( \mu \) as in the table below:
Density \( \mu=230 \) \( \mu= 259 \) \( f_1 \) .005399 .026609 \( f_2 \) .000087 .033322 \( f_3 \) .000595 .039894 - The likelihood value for the data above, assuming that these are independent samples of the true Gaussian distribution, is given by the product of the respective densities.
- By performing a numerical optimization, we can recover the “most likely” choice of \( \mu \) given our observations.
- However, for a Gaussian distribution as above, it can be demonstrated that the sample mean is the maximum likelyhood estimate of \( \mu \).

- As with the toy example before, we can apply the same method to estimating the regression parameters \( \beta_0,\beta_1 \).
- Recall that we assume \( \epsilon_i \sim N(0,\sigma^2) \), such that
- each observation is iid drawn from a Gaussian distribution, \[ Y_i \sim N(\beta_0+ \beta_1 X_i, \sigma^2) \]
- To put this in the framework of a likelihood function, we suppose that \( \overline{\beta}_0,\overline{\beta}_1,\overline{\sigma} \) are free parameters, such that \[ \begin{align}\mathcal{L}\left(\overline{\beta}_0,\overline{\beta}_1,\overline{\sigma}\vert Y_i\right)& = p_{\overline{\beta}_0,\overline{\beta}_1,\overline{\sigma}}\left(Y_i\right) \\ &=\frac{1}{\sqrt{2\pi}\overline{\sigma}}\mathrm{exp}\left\{-\frac{1}{2}\left(\frac{Y_i - \overline{\beta}_0 - \overline{\beta}_1 X_i}{\overline{\sigma}}\right)^2\right\} \end{align} \]
- Then, due to the independence, the likelihood over the set of the \( n \) observations is given simply as the product of the individual likelihoods: \[ \mathcal{L}\left(\overline{\beta}_0,\overline{\beta}_1,\overline{\sigma}\vert Y_{i=1,\cdots,n}\right) =\prod_{i=1}^n\frac{1}{\sqrt{2\pi}\overline{\sigma}}\mathrm{exp}\left\{-\frac{1}{2}\left(\frac{Y_i - \overline{\beta}_0 - \overline{\beta}_1 X_i}{\overline{\sigma}}\right)^2\right\} \]
- By maximizing the above product over all the observed data, with respect to the choice of the free parameters \( \overline{\beta}_0,\overline{\beta}_1,\overline{\sigma} \), we obtain the maximium likelihood estimate.

**Q:**Suppose we want to maximize the likelihood function, \[ \mathcal{L}\left(\overline{\beta}_0,\overline{\beta}_1,\overline{\sigma}\vert Y_{i=1,\cdots,n}\right) =\prod_{i=1}^n\frac{1}{\sqrt{2\pi}\overline{\sigma}}\mathrm{exp}\left\{-\frac{1}{2}\left(\frac{Y_i - \overline{\beta}_0 - \overline{\beta}_1 X_i}{\sigma}\right)^2\right\}; \] what approach can we take to make this tractible?- The function \( \log() \) is monotonic, so that an increase in the argument corresponds to an increase in the output.
- Using this fact, we can reduce our analysis of the likilhood function to something more simple, the “log-likelihood”, \[ \log(\mathcal{L}) = constant - n\log{\overline{\sigma}} - \frac{1}{2\overline{\sigma}^2}\sum_{i=1}^n \left(Y_i - \overline{\beta}_0 - \overline{\beta}_1 X_i \right)^2; \]
- finding the maximum value of the above equation is equivalent to finding the maximum of the likelihood, due to the monotonicity.
- Note, we write a number of values as “constant”, because for purposes of optimization in the free parameters \( \overline{\beta}_0,\overline{\beta}_1,\overline{\sigma} \), these make no difference in the outcome.
**Q:**what would be a plausible approach of finding the maximum of the above?**A:**one approach is to take the derivative in each free parameter, as before.- This will result in this case in three equations which will resemble the normal equations of the least squares approach.

- We stated that, as an addendum to the Gauss-Markov theorem, the parameter estimates by least squares \( \hat{\beta}_0,\hat{\beta}_1 \) are also the maximum likelihood parameters in the case of Gaussian errors.
- However, we note, the maximum likelihood estimate above simultaneously estimates the unknown variance \( \sigma^2 \).
- As opposed to the unbiased estimate of the variance, \[ \hat{\sigma}^2 \triangleq \frac{RSS}{n-p}, \] the maximum likelihood estimate is biased.
- Indeed, the maximum likelihood estimate for the variance is given by the naive normalization, \[ \sum_{i=1}^n \frac{\left(Y_i - \hat{Y}_i\right)^2}{n}. \]
- For large values of \( n \), the difference in the denominator between \( n \) and \( n-p \) may become small (with a reasonable number of parameters), but for small sample sizes this is quite an important difference.
- We can see that maximum likelihood estimate will severely underestimate \( \sigma^2 \) when \( n \) is small or for any \( p \) close to \( n \).

- So far, we have taken the relatively simple formulation of the regression problem in which the explanatory variable \( X \) is assumed to be a controllable constant value.
- In this context, the variation that we refer to in \( \epsilon_i \) is prototypically defined in terms of repeated sampling of replicated cases.
- Particularly, we are considering scenarios in which we can hold the value of \( X_i \) constant, and take multiple independent samples of the associated value of the response \( Y_i \).
- However, there are situations in which we want to consider
**both**the response \( Y \) and the predictor \( X \) to be random variables. - For example, an analyst may may be interested in two variables “height of person” and “weight of person” in a study of a sample of persons, with each variable being taken as random.
- The analyst might wish to study the relation between the two variables or might be interested in
- making inferences about weight of a person on the basis of the person’s height;
- making inferences about height on the basis of weight;
- or in both.

- In this context, we can treat our regression as a
**correlation model**. - In the case of a joint Gaussian distribution for \( Y \) and \( X \), our techniques in regression remain essentially the same, but the framework for our analysis and the meaning of parameters will change.
- We will discuss this change of framework in the remaining lecture.

Courtesy of: Kutner, M. et al. Applied Linear Statistical Models 5th Edition

- To begin, we will recall the basis of the Gaussian correlation model — the multivariate Gaussian distribution.
- Reframing slightly, we suppose that \( Y_1 \) and \( Y_2 = X \) are jointly distributed with respect to a Gaussian distribution as on the left.
- In this case, we will re-write \( X \) as above, as its meaning is symmetric with the response.
- The bivariate Gaussian has the functional form of the below, where
- \( \mu_i,\sigma_i \) are the mean and standard deviation of \( Y_i \) respectively;
- \( \rho_{12} \triangleq \frac{\sigma_{12}^2}{\sigma_1 \sigma_2} \) is the
**coefficient of correlation**between \( Y_1,Y_2 \); - in the figure on the left, we identify \( f(Y_1,Y_2) \) as \( p_{Y_1,Y_2}(y_1,y_2) \).

\[ \begin{align}
p_{Y_1,Y_2}(y_1,y_2) = \frac{1}{2 \pi \sigma_1 \sigma_2 \sqrt{1 - \rho_{12}^2}}\mathrm{exp}\left\{\frac{1}{2\left(1 - \rho_{12}^2\right)} \left[\left(\frac{y_1 - \mu_1}{\sigma_1}\right)^2 + \left(\frac{y_2 - \mu_2}{\sigma_2}\right)^2 - 2\rho_{12}\left( \frac{y_1 - \mu_1}{\sigma_1}\right)\left( \frac{y_2 - \mu_2}{\sigma_2}\right)\right] \right\}
\end{align} \]