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File:Confidenceinterval.png

In statistics, a confidence interval (CI) is an interval estimate of a population parameter. Instead of estimating the parameter by a single value, an interval likely to include the parameter is given. How likely the interval is to contain the parameter is determined by the confidence level or confidence coefficient. Increasing the desired confidence level will widen the confidence interval.

Confidence intervals are used to indicate the reliability of an estimate. For example, a CI can be used to describe how reliable survey results are. All other things being equal, a survey result with a small CI is more reliable than a result with a large CI.

More specifically, for a given associated proportion p (or confidence level), a CI for a population parameter is an interval that is calculated from a random sample of an underlying population such that, if the sampling was repeated numerous times and the confidence interval recalculated from each sample according to the same method, a proportion p of the confidence intervals would contain the population parameter in question. In unusual cases, a confidence set may consist of a collection several separate intervals, which may include semi-infinite intervals, and it is possible that an outcome of a confidence-interval calculation could be the set of all values from minus infinity to plus infinity.

Confidence intervals are the most prevalent form of interval estimation. Interval estimates may be contrasted with point estimates and have the advantage over these as summaries of a dataset in that more information is conveyed – not just a "best estimate" of a parameter but an indication of the accuracy with which the parameter is known.

Confidence intervals play a similar role in frequentist statistics to the credibility interval in Bayesian statistics. However, confidence intervals and credibility intervals are not only mathematically different; they have radically different interpretations.

The concept of a confidence interval for a single quantity can be generalised to be able to deal with several quantities simultaneously, in which case they are called confidence regions. Such regions can indicate not only the extent of likely estimation errors but can also reveal whether (for example) if the estimate for one quantity is too large then the other is also likely to be too large.

In modern applied practice, confidence intervals are often stated at the 95% level.[2] However, when presented graphically, confidence intervals or confidence regions may be shown for several confidence levels, for example 50%, 90% and 99%.

Theoretical basis

Definition

CI's as random intervals

Confidence intervals are constructed on the basis of a given dataset: x denotes the set of observations in the dataset, and X is used when considering the outcomes that might have been observed from the same population, where X is treated as a random variable whose observed outcome is X = x. A confidence interval is specified by a pair of functions u(.) and v(.) and the confidence interval for the given data set is defined as the interval (u(x), v(x)). To complete the definition of a confidence interval, there needs to be a clear understanding of the quantity for which the CI provides an interval estimate. Suppose this quantity is w. The property of the rules u(.) and v(.) that makes the interval (u(x),v(x)) closest to what a confidence interval for w would be, relates to the properties of the set of random intervals given by (u(X),v(X)): that is treating the end-points as random variables. This property is the coverage probability or the probability c that the random interval includes w,

c=\Pr(u(X)<w<v(X)).

Here the endpoints U = u(X) and V = v(X) are statistics (i.e., observable random variables) which are derived from values in the dataset. The random interval is (UV).

Confidence intervals for inference

For the above to provide a viable means to statistical inference, something further is required: a tie between the quantity being estimated and the probability distribution of the outcome X. Suppose that this probability distribution is characterised by the unobservable parameter θ, which is a quantity to be estimated, and by other unobservable parameters φ which are not of immediate interest. These other quantities φ in which there is no immediate interest are called nuisance parameters, as statistical theory still needs to find some way to deal with them.

The definition of a confidence interval for θ is, for a given α,

{\Pr}_{X;\theta,\phi}(u(X)<\theta<v(X))=\alpha   for all (\theta,\phi).\,

The number \alpha (sometimes reported as a percentage (100%·\alpha)) is called the confidence level or confidence coefficient. Here {\Pr}_{X;\theta,\phi} is used to indicate the probability when the random variable X has the distribution characterised by (\theta,\phi). An important part of this specification is that the random interval (UV) covers the unknown value θ with a high probability no matter what the true value of θ actually is.

Note that here {\Pr}_{X;\theta,\phi} need not refer to an explicitly given parameterised family of distributions, although it often does. Just as the random variable X notionally corresponds to other possible realisations of x from the same population or from the same version of reality, the parameters (\theta,\phi) indicate that we need to consider other versions of reality in which the distribution of X might have different characteristics.

Intervals for random outcomes

Confidence intervals can be defined for random quantities as well as for fixed quantities as in the above. See prediction interval. For this, consider an additional single-valued random variable Y which may or may not be statistically dependent on X. Then the rule for for constructing the interval(u(x), v(x)) provides a confidence interval for the as-yet-to-be observed value y of Y if

{\Pr}_{X,Y;\theta,\phi}(u(X)<Y<v(X))=\alpha   for all (\theta,\phi).\,

Here {\Pr}_{X,Y;\theta,\phi} is used to indicate the probability over the joint distribution of the random variables (X,Y) when this is characterised by parameters (\theta,\phi).

Approximate confidence intervals

For non-standard applications it is sometimes not possible to find rules for constructing confidence intervals that have exactly the required properties. But practically useful intervals can still be found. The coverage probability c(\theta,\phi) for a random interval is defined by

{\Pr}_{X;\theta,\phi}(u(X)<\theta<v(X))=c(\theta,\phi)

and the rule for constructing the interval may be accepted as providing a confidence interval if

c(\theta,\phi)\approxeq \alpha   for all (\theta,\phi)

to an acceptable level of approximation.

Comparison to Bayesian interval estimates

A Bayesian interval estimate is called a credible interval. Using much of the same notation as above, the definition of a credible interval for the unknown true value of θ is, for a given α,

{\Pr}_{\Theta|X=x}(u(x)<\Theta<v(x))=\alpha.

Here Θ is used to emhasize that the unknown value of \theta is being treated as a random variable. The definitions of the two types of intervals may be compared as follows.

  • The definition of a confidence interval involves probabilities calculated from the distribution of X for given (\theta,\phi) (or conditional on these values) and the condition needs to hold for all values of (\theta,\phi).
  • The definition of a credible interval involves probabilities calculated from the distribution of Θ conditional on the observed values of X=x and marginalised (or averaged) over the values of \Phi, where this last quantity is the random variable corresponding to the uncertainty about the nuisance parameters in \phi.

Note that the treatment of the nuisance parameters above is often omitted from discussions comparing confidence and credible intervals but it is markedly different between the two cases.

In some simple standard cases, the intervals produced as confidence and credible intervals from the same data set can be identical. They are always very different if moderate or strong prior information is included in the Bayesian analysis.

Desirable properties

When applying fairly standard statistical procedures, there will often be fairly standard ways of constructing confidence intervals. These will have been devised so as to meet certain desirable properties, which will hold given that the assumptions on which the procedure rely are true. In non-standard applications, the same desirable properties would be sought. These desirable properties may be described as: validity, optimality and invariance. Of these "validity" is most important, followed closely by "optimality". "Invariance" may be considered as a property of the method of derivation of a confidence interval rather than of the rule for constructing the interval.

  • Validity. This means that the nominal coverage probability (confidence level) of the confidence interval should hold, either exactly or to a good approximation.
  • Optimality. This means that the rule for constructing the confidence interval should make as much use of the information in the data-set as possible. Recall that one could throw away half of a dataset and still be able to derive a valid confidence interval. One way of assessing optimality is by the length of the interval, so that a rule for constructing a confidence interval is judged better than another if it leads to intervals whose widths are typically shorter.
  • Invariance. In many applications the quantity being estimated might not be tightly defined as such. For example, a survey might result in an estimate of the median income in a population, but it might equally be considered as providing an estimate of the logarithm of the median income, given that this is a common scale for presenting graphical results. It would be desirable that the method used for constructing a confidence interval for the median income would give equivalent results when applied to constructing a confidence interval for the logarithm of the median income: specifically the values at the ends of the latter interval would be the logarithms of the values at the ends of former interval.

Methods of derivation

For non-standard applications, there are several routes that might be taken to derive a rule for the construction of confidence intervals. Established rules for standard procedures might be justified or explained via several of these routes. Typically a rule for constructing confidence intervals is closely tied to a particular way of finding a point estimate of the quantity being considered.

Sample statistics
This is closely related to the method of moments for estimation. A simple example arises where the quantity to be estimated is the mean, in which case an natural estimate is the sample mean. The usual arguments indicate that the sample variance can be used to estimate the variance of the sample mean. A naive confidence interval for the true mean can be constructed centred on the sample mean with a width which is a multiple of the square root of the sample variance.
Likelihood theory
Where estimates are constructed using the maximum likelihood principle, the theory for this provides two ways of constructing confidence intervals or confidence regions for the estimates.
Estimating equations
The estimation approach here can be considered as both a generalisation of the method of moments and a generalisation of the maximum likelihood approach. There are corresponding generalisations of the results of maximum likelihood theory that allow confidence intervals to be constructed based on estimates derived from estimating equations.
Via significance testing
If significance tests are available for general values of a parameter, then confidence intervals/regions can be constructed by including in the 100p% confidence region all those points for which the significance test of the null hypothesis that the true value is the given value is not rejected at a significance level of (1-p).

Practical example

A machine fills cups with margarine, and is supposed to be adjusted so that the mean content of the cups is close to 250 grams of margarine. Of course it is not possible to fill every cup with exactly 250 grams of margarine. Hence the weight of the filling can be considered to be a random variable X. The distribution of X is assumed here to be a normal distribution with unknown expectation μ and (for the sake of simplicity) known standard deviation σ = 2.5 grams. To check if the machine is adequately adjusted, a sample of n = 25 cups of margarine is chosen at random and the cups weighed. The weights of margarine are X_1,\dots,X_{25}, a random sample from X.

To get an impression of the expectation μ, it is sufficient to give an estimate. The appropriate estimator is the sample mean:

\hat \mu=\bar X=\frac{1}{n}\sum_{i=1}^n X_i.

The sample shows actual weights x_1,\dots,x_{25}, with mean:

\bar x=\frac {1}{25} \sum_{i=1}^{25} x_i = 250.2\,\mathrm{grams}.

If we take another sample of 25 cups, we could easily expect to find values like 250.4 or 251.1 grams. A sample mean value of 280 grams however would be extremely rare if the mean content of the cups is in fact close to 250g. There is a whole interval around the observed value 250.2 of the sample mean within which, if the whole population mean actually takes a value in this range, the observed data would not be considered particularly unusual. Such an interval is called a confidence interval for the parameter μ. How do we calculate such an interval? The endpoints of the interval have to be calculated from the sample, so they are statistics, functions of the sample X_1,\dots,X_{25} and hence random variables themselves.

In our case we may determine the endpoints by considering that the sample mean \bar X from a normally distributed sample is also normally distributed, with the same expectation μ, but with standard error \sigma/\sqrt{n} = 0.5 (grams). By standardizing we get a random variable

Z = \frac {\bar X-\mu}{\sigma/\sqrt{n}} =\frac {\bar X-\mu}{0.5}

dependent on μ, but with a standard normal distribution independent of the parameter μ to be estimated. Hence it is possible to find numbers −z and z, independent of μ, where Z lies in between with probability 1 − α, a measure of how confident we want to be. We take 1 − α = 0.95. So we have:

P(-z\le Z\le z) = 1-\alpha = 0.95.

The number z follows from:

\Phi(z) = P(Z \le z) = 1 - \frac{\alpha}2 = 0.975\,,
z=\Phi^{-1}(\Phi(z)) = \Phi^{-1}(0.975) = 1.96\,,

(see probit and cumulative distribution function), and we get:

0.95 = 1-\alpha=P(-z \le Z \le z)=P \left(-1.96 \le \frac {\bar X-\mu}{\sigma/\sqrt{n}} \le 1.96 \right)
=P \left( \bar X - 1.96 \frac{\sigma}{\sqrt{n}} \le \mu \le \bar X + 1.96 \frac{\sigma}{\sqrt{n}}\right)
=P\left(\bar X - 1.96 \times 0.5 \le \mu \le \bar X + 1.96 \times 0.5\right)
=P \left( \bar X - 0.98 \le \mu \le \bar X + 0.98 \right).

This might be interpreted as: with probability 0.95 to one we will choose a confidence interval in which we will meet the parameter μ between the stochastic endpoints, but that does not mean that possibility of meeting parameter μ in confidence interval is 95% :

 \bar X - 0{.}98

and

 \bar X + 0.98.

Every time the measurements are repeated, there will be another value for the mean \bar X of the sample. In 95% of the cases μ will be between the endpoints calculated from this mean, but in 5% of the cases it will not be. The actual confidence interval is calculated by entering the measured weights in the formula. Our 0.95 confidence interval becomes:

(\bar x - 0.98;\bar x + 0.98) = (250.2 - 0.98; 250.2 + 0.98) = (249.22; 251.18).\,

This interval has fixed endpoints, where μ might be in between (or not). There is no probability of such an event. We cannot say: "with probability (1 − α) the parameter μ lies in the confidence interval." We only know that by repetition in 100(1 − α) % of the cases μ will be in the calculated interval. In 100α % of the cases however it doesn't. And unfortunately we don't know in which of the cases this happens. That's why we say: with confidence level 100(1 − α) % μ lies in the confidence interval."

The following picture shows 50 realisations of a confidence interval for μ.

Observation of the sample means we choose from the population of all realisations. There the probability is 95% we end up having chosen an interval that contains the parameter. After realisation we just have our chosen interval. As seen from the picture there was a fair chance we choose an interval containing μ; however we may be unlucky and have picked the wrong one. We'll never know; we're stuck with our interval.

Theoretical example

Suppose X1, ..., Xn are an independent sample from a normally distributed population with mean μ and variance σ2. Let

\overline{X}=(X_1+\cdots+X_n)/n,\,
S^2=\frac{1}{n-1}\sum_{i=1}^n\left(X_i-\overline{X}\,\right)^2.

Then

T=\frac{\overline{X}-\mu}{S/\sqrt{n}}

has a Student's t-distribution with n − 1 degrees of freedom. Note that the distribution of T does not depend on the values of the unobservable parameters μ and σ2; i.e., it is a pivotal quantity. If c is the 95th percentile of this distribution, then

\Pr\left(-c<T<c\right)=0.9.\,

(Note: "95th" and "0.9" are correct in the preceding expressions. There is a 5% chance that T will be less than −c and a 5% chance that it will be larger than +c. Thus, the probability that T will be between −c and +c is 90%.)

Consequently

\Pr\left(\overline{X}-cS/\sqrt{n}<\mu<\overline{X}+cS/\sqrt{n}\right)=0.9\,

and we have a theoretical (stochastic) 90% confidence interval for μ.

After observing the sample we find values \overline{x} for \overline{X} and s for S, from which we compute the confidence interval

[\overline{x}-cs/\sqrt{n},\overline{x}+cs/\sqrt{n}]\,,

an interval with fixed numbers as endpoints, of which we can no more say there is a certain probability it contains the parameter μ. Either μ is in this interval or isn't.

Meaning and interpretation

Confidence intervals in measurement

The results of measurements are often accompanied by confidence intervals. For instance, suppose a scale is known to yield the actual mass of an object plus a normally distributed random error with mean 0 and known standard deviation σ. If we weigh 100 objects of known mass on this scale and report the values ±σ, then we can expect to find that around 68% of the reported ranges include the actual mass.

If we wish to report values with a smaller standard error value, then we repeat the measurement n times and average the results. Then the 68.2% confidence interval is \pm \sigma/\sqrt{n}. For example, repeating the measurement 100 times reduces the confidence interval to 1/10 of the original width.

Note that when we report a 68.2% confidence interval (usually termed standard error) as v ± σ, this does not mean that the true mass has a 68.2% chance of being in the reported range. In fact, the true mass is either in the range or not. How can a value outside the range be said to have any chance of being in the range? Rather, our statement means that 68.2% of the ranges we report using ± σ are likely to include the true mass.

This is not just a quibble. Under the incorrect interpretation, each of the 100 measurements described above would be specifying a different range, and the true mass supposedly has a 68% chance of being in each and every range. Also, it supposedly has a 32% chance of being outside each and every range. If two of the ranges happen to be disjoint, the statements are obviously inconsistent. Say one range is 1 to 2, and the other is 2 to 3. Supposedly, the true mass has a 68% chance of being between 1 and 2, but only a 32% chance of being less than 2 or more than 3. The incorrect interpretation reads more into the statement than is meant.

On the other hand, under the correct interpretation, each and every statement we make is really true, because the statements are not about any specific range. We could report that one mass is 10.2 ± 0.1 grams, while really it is 10.6 grams, and not be lying. But if we report fewer than 1000 values and more than two of them are that far off, we will have some explaining to do.

It is also possible to estimate a confidence interval without knowing the standard deviation of the random error. This is done using the t distribution, or by using non-parametric resampling methods such as the bootstrap, which do not require that the error have a normal distribution.

How to understand confidence intervals

For users of frequentist methods, various interpretations of a confidence interval can be given.

  • The confidence interval can be expressed in terms of samples (or repeated samples): "Were this procedure to be repeated on multiple samples, the calculated confidence interval (which would differ for each sample) would encompass the true population parameter 90% of the time."
  • The explanation of a confidence interval can amount to something like: "The confidence interval represents values for the population parameter for which the difference between the parameter and the observed estimate is not statistically significant at the 10% level". In fact, this relates to one particular way in which a confidence interval may be constructed.
  • The probability associated with a confidence interval may also be considered from a pre-experiment point of view, in the same context in which arguments for the random allocation of treatments to study items are made. Here the experimenter sets out the way in which they intend to calculate a confidence interval and know, before they do the actual experiment, that the interval they will end up calculating has a certain chance of covering the true but unknown value. This is very similar to the "repeated sample" interpretation above, except that it avoids relying on considering hypothetical repeats of a sampling procedure that may not be repeatable in any meaningful sense.

Critics of frequentist methods dislike at least the first two of the three interpretations given above. Users of Bayesian methods, if they produced an interval estimate, would by contrast want to say "My degree of belief that the parameter is in fact in this interval is 90%". Disagreements about these issues are not disagreements about solutions to mathematical problems. Rather they are disagreements about the ways in which mathematics is to be applied.



Confidence levels are typically given alongside statistics resulting from sampling.

In a statement "we are 90% confident that between 35% and 45% of voters favor Candidate A", 90% is our confidence level and 35%-45% is our confidence interval.

It is very tempting to misunderstand this statement in the following way. We used capital letters U and V for random variables; it is conventional to use lower-case letters u and v for their observed values in a particular instance. The misunderstanding is the conclusion that[How to reference and link to summary or text]

\Pr(u<\theta<v)=0.9,\,

so that after the data has been observed, a conditional probability distribution of θ, given the data, is inferred. For example, suppose X is normally distributed with expected value θ and variance 1. (It is grossly unrealistic to take the variance to be known while the expected value must be inferred from the data, but it makes the example simple.) The random variable X is observable. (The random variable X − θ is not observable, since its value depends on θ.) Then X − θ is normally distributed with expectation 0 and variance 1. Given that 90% of the standard normal distribution lies between −1.645 and 1.645, we know:

\Pr(-1.645<X-\theta<1.645)=0.9.\,

Consequently

\Pr(X-1.645<\theta<X+1.645)=0.9,\,

so the interval from X − 1.645 to X + 1.645 is a 90% confidence interval for θ. But when X = 82 is observed, can we then say that

\Pr(82-1.645<\theta<82+1.645)=0.9\  \mbox{?}

This conclusion does not follow from the laws of probability because θ is not a "random variable"; i.e., no probability distribution has been assigned to it.[How to reference and link to summary or text] Confidence intervals are generally a frequentist method, i.e., employed by those who interpret "90% probability" as "occurring in 90% of all cases".[How to reference and link to summary or text] Suppose, for example, that θ is the mass of the planet Neptune, and the randomness in our measurement error means that 90% of the time our statement that the mass is between this number and that number will be correct. The mass is not what is random. Therefore, given that we have measured it to be 82 units, we cannot say that in 90% of all cases, the mass is between 82 − 1.645 and 82 + 1.645. There are no such cases; there is, after all, only one planet Neptune.

But if probabilities are construed as degrees of belief rather than as relative frequencies of occurrence of random events, i.e., if we are Bayesians rather than frequentists, can we then say we are 90% sure that the mass is between 82 − 1.645 and 82 + 1.645? Many answers to this question have been proposed, and are philosophically controversial. The answer will not be a mathematical theorem, but a philosophical tenet. Less controversial are Bayesian credible intervals, in which one starts with a prior probability distribution of θ, and finds a posterior probability distribution, which is the conditional probability distribution of θ given the data.

Robust confidence intervals

In the process of weighing 1000 objects, under practical conditions, it is easy to believe that the operator might make a mistake in procedure and so report an incorrect mass (thereby making one type of systematic error). Suppose he has 100 objects and he weighed them all, one at a time, and repeated the whole process ten times. Then he can calculate a sample standard deviation for each object, and look for outliers. Any object with an unusually large standard deviation probably has an outlier in its data. These can be removed by various non-parametric techniques. If he repeated the process only three times, he would simply take the median of the three measurements and use σ to give a confidence interval. The 200 extra weighings served only to detect and correct for operator error and did nothing to improve the confidence interval. With more repetitions, he could use a truncated mean, discarding say the largest and smallest values and averaging the rest. He could then use a bootstrap calculation to determine a confidence interval narrower than that calculated from σ, and so obtain some benefit from a large amount of extra work.

These procedures are robust against procedural errors which are not modeled by the assumption that the balance has a fixed known standard deviation σ. In practical applications where the occasional operator error can occur, or the balance can malfunction, the assumptions behind simple statistical calculations cannot be taken for granted. Before trusting the results of 100 objects weighed just three times each to have confidence intervals calculated from σ, it is necessary to test for and remove a reasonable number of outliers (testing the assumption that the operator is careful and correcting for the fact that he is not perfect), and to test the assumption that the data really have a normal distribution with standard deviation σ.

The theoretical analysis of such an experiment is complicated, but it is easy to set up a spreadsheet which draws random numbers from a normal distribution with standard deviation σ to simulate the situation (use =norminv(rand(),0,σ)). See for example Wittwer, J.W., "Monte Carlo Simulation in Excel: A Practical Guide", June 1, 2004. These techniques also work in Open Office and gnumeric.

After removing obvious outliers, one could subtract the median from the other two values for each object, and examine the distribution of the 200 resulting numbers. It should be normal with mean near zero and standard deviation a little larger than σ. A simple Monte Carlo spreadsheet calculation would reveal typical values for the standard deviation (around 105 to 115% of σ). Or, one could subtract the mean of each triplet from the values, and examine the distribution of 300 values. The mean is identically zero, but the standard deviation should be somewhat smaller (around 75 to 85% of σ).

Confidence intervals for proportions and related quantities

See also: Margin of error

An approximate confidence interval for a population mean can be constructed for random variables that are not normally distributed in the population, relying on the central limit theorem, if the sample sizes and counts are big enough. The formulae are identical to the case above (where the sample mean is actually normally distributed about the population mean). The approximation will be quite good with only a few dozen observations in the sample if the probability distribution of the random variable is not too different from the normal distribution (e.g. its cumulative distribution function does not have any discontinuities and its skewness is moderate).

One type of sample mean is the mean of an indicator variable, which takes on the value 1 for true and the value 0 for false. (Statisticians often call indicator variables "dummy variables", but that term is also frequently used by mathematicians for the concept of a bound variable.) The mean of such a variable is equal to the proportion that have the variable equal to one (both in the population and in any sample). Thus, the sample mean for a variable labeled MALE in data is just the proportion of sampled observations who have MALE = 1, i.e. the proportion who are male. This is a useful property of indicator variables, especially for hypothesis testing.

To apply the central limit theorem, one must use a large enough sample. A rough rule of thumb is that one should see at least 5 cases in which the indicator is 1 and at least 5 in which it is 0. Confidence intervals constructed using the above formulae may include negative numbers or numbers greater than 1, but proportions obviously cannot be negative or exceed 1. The probability assigned to negative numbers and numbers greater than 1 is usually small when the sample size is large and the proportion being estimated is not too close to 0 or 1.

Confidence intervals for cases where the method above assigns a substantial probability to (−∞, 0) or to (1, ∞) may be constructed by inverting hypothesis tests. If we think of conducting hypothesis tests over the whole feasible range of parameter values, and including any values for which a single hypothesis test would not reject the null hypothesis that the true value was that value, given our sample value, we can make a confidence interval based on the central limit theorem that does not violate the basic properties of proportions.

On the other hand, sample proportions can only take on a finite number of values, so the central limit theorem and the normal distribution are not the best tools for building a confidence interval. A better method would rely on the binomial distribution or the beta distribution, and there are a number of better methods in widespread use. For details on advantages and disadvantages of each, see:

  • "Interval Estimation for a Binomial Proportion", Lawrence D. Brown, T. Tony Cai, Anirban DasGupta, Statistical Science, volume 16, number 2 (May, 2001), pages 101-117.

See also

Online calculators

References

  1. Goldstein, H., & Healey, M.J.R. (1995). "The graphical presentation of a collection of means." Journal of the Royal Statistical Society, 158, 175-77.
  2. Zar, J.H. (1984) Biostatistical Analysis. Prentice Hall International, New Jersey. pp 43-45


Further reading

  • Abelson, R. P. (1997). On the surprising longevity of flogged horses: Why there is a case for the significance test: Psychological Science Vol 8(1) Jan 1997, 12-15.
  • Aiken, L. R. (1983). An interactive program for testing the significance of various correlation coefficients: Educational and Psychological Measurement Vol 43(1) Spr 1983, 181-182.
  • Alf, E. F., Jr., & Graf, R. G. (1999). Asymptotic confidence limits for the difference between two squared multiple correlations: A simplified approach: Psychological Methods Vol 4(1) Mar 1999, 70-75.
  • Alf, E. F., & Grossberg, J. M. (1979). The geometric mean: Confidence limits and significance tests: Perception & Psychophysics Vol 26(5) Nov 1979, 419-421.
  • Algina, J. (1999). A comparison of methods for constructing confidence intervals for the squared multiple correlation coefficient: Multivariate Behavioral Research Vol 34(4) 1999, 493-504.
  • Algina, J., & Keselman, H. J. (2003). Approximate confidence intervals for effect sizes: Educational and Psychological Measurement Vol 63(4) Aug 2003, 537-553.
  • Algina, J., Keselman, H. J., & Penfield, R. D. (2006). Confidence Interval Coverage for Cohen's Effect Size Statistic: Educational and Psychological Measurement Vol 66(6) Dec 2006, 945-960.
  • Algina, J., & Moulder, B. C. (2001). Sample sizes for confidence intervals on the increase in the squared multiple correlation coefficient: Educational and Psychological Measurement Vol 61(4) Aug 2001, 633-649.
  • Anderson, A. J. (2003). Utility of a dynamic termination criterion in the ZEST adaptive threshold method: Vision Research Vol 43(2) Jan 2003, 165-170.
  • Andrade, A. G., & Scalco, M. Z. (1989). Concepts of validity and confidence in an instrument for evaluating treatment for alcoholics: Jornal Brasileiro de Psiquiatria Vol 38(1) Jan-Feb 1989, 41-45.
  • Aoshima, M., & Mukhopadhyay, N. (1998). Fixed-Width Simultaneous Confidence Intervals for Multinormal Means in Several Intraclass Correlation Models: Journal of Multivariate Analysis Vol 66(1) Jul 1998, 46-63.
  • Appel, V., & Kipnis, D. (1954). The use of levels of confidence in item analysis: Journal of Applied Psychology Vol 38(4) Aug 1954, 256-259.
  • Atkinson, L. (1990). Standard errors of prediction for the Vineland Adaptive Behavior Scales: Journal of School Psychology Vol 28(4) Win 1990, 355-359.
  • Attorresi, H. F., Aguerri, M. E., Lozzia, G. S., & Galibert, M. S. (2004). Confidence intervals or true scores. Their assumptions made explicit: Interdisciplinaria Revista de Psicologia y Ciencias Afines Vol 21(1) 2004, 29-51.
  • Baranski, J. V., & Petrusic, W. M. (1994). The calibration and resolution of confidence in perceptual judgments: Perception & Psychophysics Vol 55(4) Apr 1994, 412-428.
  • Barchard, K. A., & Hakstian, A. R. (1997). The robustness of confidence intervals for coefficient alpha under violation of the assumption of essential parallelism: Multivariate Behavioral Research Vol 32(2) 1997, 169-191.
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