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Examples in this article were generated with
4.4.1
by the package PowerTOST
.^{1}
Have you ever wondered where all these nice numbers come from?
\(\small{\alpha\;0.05}\), bioequivalence limits \(\small{80.00125.00\%}\), switching to SABE at \(\small{CV_\text{wR}\;30\%}\), \(\small{k\;0.760}\), upper cap of scaling at \(\small{CV_\text{wR}\;50\%}\) or \(\small{57.4\%}\), \(\small{s_\text{wR}\;0.294}\), you name it.
The list is endless and as interesting as a telephone directory.
Why this number? OK, at least this one is easy to explain: Mr. Fisher considered it convenient.
“The value for which p = 0.05, or 1 in 20, is 1.96 or nearly 2; it is convenient to take this point as a limit in judging whether a deviation is to be considered significant or not.
For nearly a century this number is the holy grail of frequentists. Members of the other chuch, the Bayesians, hold other believes.
Did you ever wonder why (based on Mr Fisher’s \(\small{\alpha=0.05}\)) we use a \(\small{100\,(12\,\alpha)=90\%}\) confidence interval (CI) in bioequivalence (BE) and not the \(\small{95\%}\) CI like in clinical Phase III?
In Phase III we want to demonstrate superiority (for nitpickers: noninferiority) of verum to placebo. Once a drug is approved, we can be ≥95% sure that it is safe and efficacious (though there is still a <5% chance that even a blockbuster does not perform better than snake oil).
In the early days of BE a \(\small{95\%}\) CI was applied indeed. Below a scan of my conference proceedings.^{3}
So why did we change to the \(\small{90\%}\) CI?
In a particular patient the bioavailability can beeither too low \(\small{(p(\text{BA}<\phantom{1}80\%)>5\%)}\)
or too high \((\small{p(\text{BA}>125\%)>5\%})\)
but evidently not at the same time.
The \(\small{90\%}\)
CI controls the risk for the
population of patients. Therefore, if a study passes, the risk
for patients does still not exceed \(\small{5\%}\).^{4} ^{5} ^{6}
< 0.15
sd < 501
n < c(0, dnorm(x = 0, sd = sd))
ylim dev.new(width = 4.5, height = 2.25, record = TRUE)
< par(no.readonly = TRUE)
op par(mar = c(4, 1, 0, 0), cex.axis = 0.9)
# TOST (left)
plot(log(c(0.5, 2)), c(0, 0), type = "n", xlim = log(c(0.5, 2)), axes = FALSE,
ylim = ylim, xlab = expression(mu[T]/mu[R]*" (%)"), ylab = "")
axis(1, at = c(log(0.5), log(1/1.5), log(0.8), 0, log(1.25), log(1.5), log(2)),
labels = sprintf("%.0f", 100*c(0.5, 1/1.5, 0.8, 1, 1.25, 1.5, 2)))
mtext(side = 2, text = expression(italic(p)*" (%)"), line = 0.1)
< seq(log(0.8), log(2), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#90EE90", border = "#90EE90")
text(0, ylim[2]/2, adj = 0.5, labels = 95, cex = 0.9)
< seq(log(0.5), log(0.8), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#FA8072", border = "#FA8072")
text(log(0.8), dnorm(x = log(1.25), sd = sd)*0.2, pos = 2, offset = 0.5,
labels = 5, cex = 0.9)
# TOST (right)
plot(log(c(0.5, 2)), c(0, 0), type = "n", xlim = log(c(0.5, 2)), axes = FALSE,
ylim = ylim, xlab = expression(mu[T]/mu[R]*" (%)"), ylab = "")
axis(1, at = c(log(0.5), log(1/1.5), log(0.8), 0, log(1.25), log(1.5), log(2)),
labels = sprintf("%.0f", 100*c(0.5, 1/1.5, 0.8, 1, 1.25, 1.5, 2)))
mtext(side = 2, text = expression(italic(p)*" (%)"), line = 0.1)
< seq(log(0.5), log(1.25), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#90EE90", border = "#90EE90")
text(0, ylim[2]/2, adj = 0.5, labels = 95, cex = 0.9)
< seq(log(1.25), log(2), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#FA8072", border = "#FA8072")
text(log(1.25), dnorm(x = log(1.25), sd = sd)*0.2, pos = 4, offset = 0.5,
labels = 5, cex = 0.9)
# 90% CI inclusion
plot(log(c(0.5, 2)), c(0, 0), type = "n", xlim = log(c(0.5, 2)), axes = FALSE,
ylim = ylim, xlab = expression(mu[T]/mu[R]*" (%)"), ylab = "")
axis(1, at = c(log(0.5), log(1/1.5), log(0.8), 0, log(1.25), log(1.5), log(2)),
labels = sprintf("%.0f", 100*c(0.5, 1/1.5, 0.8, 1, 1.25, 1.5, 2)))
mtext(side = 2, text = expression(italic(p)*" (%)"), line = 0.1)
< seq(log(0.8), log(1.25), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#90EE90", border = "#90EE90")
text(0, ylim[2]/2, adj = 0.5, labels = 90, cex = 0.9)
< seq(log(0.5), log(0.8), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#FA8072", border = "#FA8072")
text(log(0.8), dnorm(x = log(1.25), sd = sd)*0.2, pos = 2, offset = 0.5,
labels = 5, cex = 0.9)
< seq(log(1.25), log(2), length.out = n)
x polygon(x = c(x, rev(x)), y = c(rep(0, n), rev(dnorm(x = x, sd = sd))),
col = "#FA8072", border = "#FA8072")
text(log(1.25), dnorm(x = log(1.25), sd = sd)*0.2, pos = 4, offset = 0.5,
labels = 5, cex = 0.9)
par(op)
If we would have kept the \(\small{95\%}\) CI, the patient’s risk would be only \(\small{2.5\%}\). Fine in principle but we are only \(\small{95\%}\) confident that the reference product works at all (see above). Hence, keeping the \(\small{95\%}\) CI would have been double standards.
In the early days of bioequivalence, the analysis was performed on raw (untransformed) data, _i.e., assuming an additive model. Based on a clinically not relevant difference \(\small{\Delta=20\%}\)^{7} one gets: \[\left\{\theta_1,\theta_2\right\}=\left\{100\,(1\Delta),100\,(1+\Delta)\right\}=80120\%\tag{1}\] One should not forget that bioequivalence was never a scientific theory in the Popperian sense but an ad hoc solution to a pressing problem in the 1970s.^{8} ^{9} The commonly assumed clinically not relevant difference \(\small{\Delta=20\%}\) is arbitrary (as any other). However, we have decades of empiric evidence that the concept is sufficient in practice. Apart from occasional anecdotal reports (mainly dealing with NTIDs), no problems are evident switching between the originator and generics (and vice versa) in terms of lacking efficacy and safety problems.
Some argued that concentrations and derived
pharmacokinetic metrics like the Area Under the Curve (AUC) do
not follow a normal distribution – which is a prerequiste of the
ANOVA. Makes sense.
If one works with the arithmetic mean and its
variance, it implies that there is a certain probability of
negative values because the domain of
\(\small{\mathcal{N}(\mu;\sigma^2)}\)
is \(\small{[\infty<x+\infty]}\)
for \(\small{x\in \mathbb{R}}\).
A captivating example of the FDA:^{10}
Splendid, a line
chart instead of a scatter
plot.
Oh dear, someone clicked the first
(in Excelparlance a line plot
instead of an
XY plot
)! Therefore, one hour intervals in the early part
of the profile appear as wide as the twelve hours at the end. Yep, in a
line chart the xvalues are ordered
and the axis is equally spaced. That gives the false impression
that the drug follows zeroorder elimination.
Any statistic implies an underlying distribution (e.g., \(\small{\overline{x},\,\text{s}\;\mapsto}\) normal)!
Did these folks really believe that at seven hours there’s a ≈16%
probability that concentrations are ≤–232 ng/mL‽
Which cult of Pastafarianism
do they adhere to? The one holding that negative mass
exist or the other believing in negative lengths?
Exploring large data sets of concentrations (and derived PK metrics like AUC and C_{max}) we see that their distributions are skewed to the right. The lognormal distribution does not only pull the right tail in but also with its domain of \(\small{]0<x+\infty]}\) for \(\small{x\in \mathbb{R}^{+}}\) avoids the ‘problem’ of negative concentrations. Its location parameter is the geometric mean.
This plot reflects the extreme variability of the drug much better and shows that at any given time point high concentrations are more likely than low ones. Of course, we see also that elimination is firstorder (exponentially decreasing).
“After a dose we know only one thing for sure: The concentration is not zero.
His statement ended in shouting matches. Don’t know why.^{12}
<nitpick></nitpick>
Another argument is based on the fundamental equation of pharmacokinetics, namely \[AUC=\frac{f\cdot D}{CL},\tag{2}\] where \(\small{f}\) is the fraction absorbed, \(\small{D}\) the dose, and \(\small{CL}\) the clearance.
If we want to compare the bioavailabilities of two drugs (\(\small{f_\text{T},f_\text{R}}\)) we arrive at \[\frac{f_\text{T}}{f_\text{R}}=\frac{AUC_\text{T}\cdot CL}{D_\text{T}}\Big{/}\frac{AUC_\text{R}\cdot CL}{D_\text{R}}.\tag{3}\] By assuming^{13} identical doses (\(\small{D_\text{T}\equiv D_\text{R}}\)) – and no interoccasion variability of clearances (\(\small{CL=\text{const}}\)) in a crossover study – we can cancel both out \[\require{cancel}\frac{f_\text{T}}{f_\text{R}}=\frac{AUC_\text{T}\cdot \cancel{CL}}{\cancel{D_\text{T}}}\Big{/}\frac{AUC_\text{R}\cdot \cancel{CL}}{\cancel{D_\text{R}}}\tag{4}\] to get what we want: \[\frac{f_\text{T}}{f_\text{R}}=\frac{AUC_\text{T}}{AUC_\text{R}}.\tag{5}\] Hey, that’s a ratio and hence, we have to use a multiplicative model.
But we need differences in the ANOVA (which is an additive model)… Nothing easier than that. \(\small{(2)}\) can be rewritten as \[\log_{e}AUC=\log_{e}f\,+\log_{e}D\,\log_{e}CL.\tag{6}\] After the same substitutions and cancelations as in \(\small{(3)}\) and \(\small{(4)}\), we can perform the analysis with \[\log_{e}PE=\log_{e}f_T\,\log_{e}f_T=\log_{e}AUC_\text{T}\,\log_{e}AUC_\text{R},\tag{7}\] and use \(\small{\exp(\log_{e}PE)}\) to get with \(\small{PE}\) a number clinicians can comprehend.^{14}
At the first BioInternational conference^{15} there was a poll among the participants about the transformation. Outcome: ⅓ never, ⅓ always, ⅓ case by case (i.e., perform both analyses and report the one with narrower CI ‘because it fits the data better’). Let’s be silent about the last camp.^{16}
Wait a minute! The original acceptance range
\(\small{(1)}\)
was symmetrical around \(\small{100\%}\). In \(\small{\log_{e}}\)scale it should be
symmetrical around \(\small{0}\)
(because \(\small{\log_{e}1=0}\)).
What happens to our \(\small{\Delta}\), which should still be
\(\small{20\%}\)? Due to the positive
skewness of the
lognormal distribution a [heated debate] lively
discussion started after early publications proposing \(\small{80125\%}\).^{17} ^{18} Keeping \(\small{80120\%}\) would have been flawed
because the maximum power should be obtained at \(\small{\mu_\text{T}/\mu_\text{R}=1}\) for
\[\exp\left((\log_{e}\theta_1+\log_{e}\theta_2)/2\right),\tag{8}\]
which works only if \(\small{\theta_2=\theta_1^{1}}\) or \(\small{\theta_1=\theta_2^{1}}\). Keeping
the original limits, maximum power would be obtained at \(\small{\mu_\text{T}/\mu_\text{R}=}\) \(\small{\exp((\log_{e}0.8+\log_{e}1.2)/2)\approx0.979796}\).
There were three parties (all agreed that the acceptance range should be symmetrical in \(\small{\log_{e}}\)scale and consequently asymmetrical when backtransformed). These were their arguments and suggestions:
\[\left\{\theta_1,\theta_2\right\}=\left\{100\,(1\Delta),100/(1\Delta)\right\}=80125\%\tag{11}\]
We all know which party prevailed. Can you imagine why?
How I love rounding the confidence interval according to the
guidelines!
All jurisdictions require the
CI in percent rounded to two
decimal places, whereas Health Canada to only one.
If we do that^{19} it will increase the patient’s risk. Why? Simple: If we would use the exact result, a study with a lower Confidence Limit (CL) of \(\small{79.995\%}\) would fail as one would with an upper CL of say, \(\small{125.005\%}\). Both would pass after rounding (for details see the Interlude below).
library(PowerTOST) # attach it to run the examples
# Cave: long runtime
# Exact rounding according to IEEE 754
< 0.95 # assumed T/Rratio
theta0 < 0.25 # assumed CV
CV0 < 0.80 # target power
target < "2x2x2"
design < 1e5L # number of simulations
nsims set.seed(123456)
< sampleN.TOST(CV = CV0, theta0 = theta0, targetpower = target,
n design = design, print = FALSE)[["Sample size"]]
< mse2CV(CV2mse(CV0) * rchisq(nsims, df = n  2) / (n  2))
CV < exp(rnorm(nsims, mean = log(theta0),
pe sd = sqrt(0.5 / n) * sqrt(CV)))
< data.frame(CV = CV, PE = 100 * pe, lower = NA_real_,
comp upper = NA_real_, exact = FALSE, round2 = FALSE,
round1 = FALSE, round0 = FALSE)
for (i in 1:nsims) {
3:4] < 100*CI.BE(CV = CV[i], pe = pe[i], n = n)
comp[i, if (comp$lower[i] >= 80 & comp$upper[i] <= 125) comp$exact[i] < TRUE
if (round(comp$lower[i], 2) >= 80 & round(comp$upper[i], 2) <= 125)
$round2[i] < TRUE
compif (round(comp$lower[i], 1) >= 80 & round(comp$upper[i], 1) <= 125)
$round1[i] < TRUE
compif (round(comp$lower[i], 0) >= 80 & round(comp$upper[i], 0) <= 125)
$round0[i] < TRUE
comp
}cat("Design :", design,
"\nCV :", CV0,
"\ntheta0 :", theta0,
"\ntarget power:", target,
"\nn :", n,
"\nPercentage of", formatC(nsims, format = "d", big.mark = ","),
"simulated studies passing\nthe acceptance limits with the 90% CI:",
"\n full precision:",
sprintf("%.3f%% (scientific)", 100 * sum(comp$exact) / nsims),
"\n 80.00 \u2013 125.00:",
sprintf("%.3f%% (guidelines)",
100 * sum(comp$round2) / nsims),
"\n 80.0 \u2013 125.0 :",
sprintf("%.3f%% (Health Canada)",
100 * sum(comp$round1) / nsims),
"\n 80 \u2013 125 :",
sprintf("%.3f%% (only for comparison)",
100 * sum(comp$round0) / nsims), "\n")
# Design : 2x2x2
# CV : 0.25
# theta0 : 0.95
# target power: 0.8
# n : 28
# Percentage of 100,000 simulated studies passing
# the acceptance limits with the 90% CI:
# full precision: 80.341% (scientific)
# 80.00 – 125.00: 80.375% (guidelines)
# 80.0 – 125.0 : 80.605% (Health Canada)
# 80 – 125 : 82.812% (only for comparison)
Do you see a pattern? Am I picky?
In all textbooks and publications the
CI inclusion approach is given
as \[\begin{matrix}\tag{12}
\theta_1=1\Delta,\theta_2=\left(1\Delta\right)^{1}\\
H_0:\;\frac{\mu_\text{T}}{\mu_\text{R}}\not\subset\left\{\theta_1,\,\theta_2\right\}\;vs\;H_1:\;\theta_1<\frac{\mu_\text{T}}{\mu_\text{R}}<\theta_2,
\end{matrix}\] where \(\small{H_0}\) is the null
hypothesis of inequivalence and \(\small{H_1}\) the alternative
hypothesis. \(\small{\theta_1}\)
and \(\small{\theta_2}\) are the lower
and upper limits of the acceptance range, and \(\small{\mu_\text{T}}\) and \(\small{\mu_\text{R}}\) the geometric least
squares means of \(\small{\text{T}}\)
and \(\small{\text{R}}\),
respectively.
Alternatively by the TOST
procedure:^{4} \[\begin{matrix}\tag{13}
H_{01}:\frac{\mu_\text{T}}{\mu_\text{R}}\leq \theta_1 & vs &
H_{11}:\frac{\mu_\text{T}}{\mu_\text{R}}>\theta_1\\
H_{02}:\frac{\mu_\text{T}}{\mu_\text{R}}\geq \theta_2 & vs &
H_{12}:\frac{\mu_\text{T}}{\mu_\text{R}}<\theta_2
\end{matrix}\] Do you see anything about rounding here?
I don’t. If you can point me to a publication supporting this regulatory
‘invention’, please let me know.
Average bioequivalence shown for confidence=0.90 and percent=0.20
Failed to show average bioequivalence for confidence=0.90 and percent=0.20
For good reasons Certara removed the text from the output in later releases.
Interlude
If you insist in rounding (I hope, you don’t), do it at least
correctly, i.e., according to the IEEE Standard for
FloatingPoint Arithmetic. That’s also called ‘Round to
Even’, where zero is considered an even number.
Let’s call the digit at the desired decimal place
x
and the digit after the desired decimal place
y
. Rules:
y
is 0, 1, 2, 3, 4 →x
down.y
is 5 (follow by other digits), 6, 7, 8, 9 →x
up.y
is 5 →x
in such a way that it becomes an even number.125.004
→ 125.00
(rule 1)125.0051
→ 125.01
(rule 2)125.005
→ 125.00
(rule 3)
That’s not necessarily the default in some software and in others it is not implemented at all. Instead, rounding ‘Away from Zero’ (commercial rounding, as we are taught at school) is used.
Let’s see what happens to 125.005
(according to rule 3
125.00
is correct, whereas 125.01
is
wrong).
Software  Function  Result  Note 

Excel 
ROUND(125.005, 2)

125.01

All versions 
MROUND(125.005, 0.01)

125.01

All versions  
Other spreadsheets 
ROUND(125.005, 2)

125.01

All versions 
Phoenix WinNonlin 
round(125.005, 2)

125.01

All versions 
JavaScript 
Math.round(125.005 * 100) / 100

125.01

All versions 
GNU Octave 
round(125.005 * 100) / 100

125.01

All versions 
PHP < 5.3.0 
round(125.005, 2)

125.01

Only one implemented 
PHP ≥ 5.3.0 
round(125.005, 2)

125

New default 
round(125.005, 2, PHP_ROUND_HALF_EVEN)

125

Preferrable for clarity  
round(125.005, 2, PHP_ROUND_HALF_UP)

125.01

Backcompatability  
SAS 
ROUND(125.005, 0.01)

125.01


ROUNDE(125.005, 0.01)

125


C 
round(125.005 * 100) / 100

125.01

∗ 
lround(125.005 * 100) / 100

125.01

∗  
printf("%.2f", 125.005)

125.00

∗  
C++ 
round(125.005 * 100) / 100

125.01

∗ 
lround(125.005 * 100) / 100

125

∗  
Java 
Math.round(125.005 * 100) / 100

125


Perl 
printf("%.2f", 125.005)

125.00


SQLite 
SELECT round(125.005, 2)

125.0


WolframAlpha 
round(125.005, 0.01)

125


Maxima 
round(125.005 * 100) / 100

125


R 
round(125.005, 2)

125

All versions 
julia 
round(125.005; digits = 2, base = 10)

125.0

All versions 
Python 
round(125.005, 2)

125.0

All versions 
∗ Compiler and variable declarations not shown 
For correct rounding in spreadsheets use
ROUND(125.005,4(1+INT(LOG10(ABS(125.005)))))
. In other
software (Phoenix WinNonlin, JavaScript, GNU Octave, C) you have to roll
out your own functions as well.
If you want to get – for comparison purposes – commercial rounding in
R
:
< function(x, y) sign(x) * trunc(abs(x) * 10^y + 0.5) / 10^y
c.round c.round(125.005, 2)
# [1] 125.01
In other IEEE 754compliant software (Java, Perl, SQLite, WolframAlpha, Maxima, julia, Python) you have also to roll out similar functions for commercial rounding.
After rounding a ‘worst case’ confidence interval, the width of the acceptance range will appear as the expected 45% but actually it is wider. The clinically relevant difference \(\small{\Delta=20\%}\) is not kept (the ‘implied’ limits are the maximum possible ones for rounding, i.e., 79.995 – 125.005%).
< AR.new < setNames(c(80, 125), c("lower", "upper"))
AR # True acceptance limits due to rounding
1:2] < c(79.995, 125.005)
AR.new[< round(AR.new, 2)
rd < BE.rd < "fail"
BE.CI if (AR.new[1] >= AR[1] & AR.new[2] <= AR[2]) BE.CI < "pass"
if (rd[1] >= AR[1] & rd[2] <= AR[2]) BE.rd < "pass"
# Examples
< seq(0.15, 0.3, 0.025)
CV < res2 < data.frame(CV = CV, n = NA_integer_,
res1 lower = NA_real_, upper = NA_real_)
# Use the defaults: design = 2x2x2, theta0 = 0.95, targetpower = 0.8
for (i in seq_along(CV)) {
$n[i] < sampleN.TOST(CV = CV[i], print = FALSE)[["Sample size"]]
res1$lower[i] < power.TOST(CV = CV[i], n = res1$n[i], theta0 = 0.80,
res1theta1 = AR.new[["lower"]] / 100,
theta2 = AR.new[["upper"]] / 100)
$upper[i] < power.TOST(CV = CV[i], n = res1$n[i], theta0 = 1.25,
res1theta1 = AR.new[["lower"]] / 100,
theta2 = AR.new[["upper"]] / 100)
$n[i] < sampleN.TOST(CV = CV[i],
res2theta1 = AR.new[["lower"]] / 100,
theta2 = AR.new[["upper"]] / 100,
print = FALSE)[["Sample size"]]
$lower[i] < power.TOST(CV = CV[i], n = res2$n[i],
res2theta0 = AR.new[["lower"]] / 100,
theta1 = AR.new[["lower"]] / 100,
theta2 = AR.new[["upper"]] / 100)
$upper[i] < power.TOST(CV = CV[i], n = res2$n[i],
res2theta0 = AR.new[["lower"]] / 100,
theta1 = AR.new[["lower"]] / 100,
theta2 = AR.new[["upper"]] / 100)
}cat("Acceptance range:",
paste(sprintf("%.2f%%", AR), collapse = " , "),
sprintf(" (midpoint: %9.5f%%)", sqrt(prod(AR))),
"\nTrue AR :",
paste(sprintf("%.3f%%", AR.new), collapse = ", "),
sprintf("(midpoint: %9.5f%%)", sqrt(prod(AR.new))),
"\nRounded AR :",
paste(sprintf("%5.2f%%", rd), collapse = " , "),
"\n\nExpected width of acceptance range:",
sprintf("%.2f%%", diff(AR)),
sprintf("(Delta %.3f%%)", 100  AR[1]),
"\nTrue width of acceptance range :",
sprintf("%.2f%%", diff(AR.new)),
sprintf("(Delta %.4f to %.3f%%)",
100 * (1  100 / AR.new[2]), 100  AR.new[1]),
"\n\nBE based on extreme CI :", BE.CI,
"\nBE based rounded extreme CI:", BE.rd)
cat("\n\nType I Errors at \u2013 exactly \u2013 80 and 125% (Delta 20%)\n")
print(round(res1, 7), row.names = FALSE)
cat("Type I Errors at the \u2018implied\u2019 limits 79.995 and 125.005%\n")
print(round(res2, 7), row.names = FALSE)
# Acceptance range: 80.00% , 125.00% (midpoint: 100.00000%)
# True AR : 79.995%, 125.005% (midpoint: 99.99887%)
# Rounded AR : 80.00% , 125.00%
#
# Expected width of acceptance range: 45.00% (Delta 20.000%)
# True width of acceptance range : 45.01% (Delta 20.0032 to 20.005%)
#
# BE based on extreme CI : fail
# BE based rounded extreme CI: pass
#
# Type I Errors at – exactly – 80 and 125% (Delta 20%)
# CV n lower upper
# 0.150 12 0.0500989 0.0500632
# 0.175 16 0.0501000 0.0500640
# 0.200 20 0.0500991 0.0500634
# 0.225 24 0.0500973 0.0500622
# 0.250 28 0.0500951 0.0500607
# 0.275 34 0.0500962 0.0500615
# 0.300 40 0.0500963 0.0500615
# Type I Errors at the ‘implied’ limits 79.995 and 125.005%
# CV n lower upper
# 0.150 12 0.0499998 0.0499998
# 0.175 16 0.0499999 0.0499999
# 0.200 20 0.0499999 0.0499999
# 0.225 24 0.0499998 0.0499998
# 0.250 28 0.0499996 0.0499996
# 0.275 34 0.0499997 0.0499997
# 0.300 40 0.0499998 0.0499998
Therefore, with rounding the probability of passing BE increases and will lead to an inflated Type I Error. Note that we see different Type I Errors at \(\small{\left\{\theta_1,\,\theta_2\right\}}\) because the ‘implied’ limits are not symmetrical in logscale.
Hence, the assessment for BE should always be performed with the CI in full numeric precision. Only then the Type I Error is strictly controlled.
These limits are stated in most guideline when dealing with Narrow Therapeutic Index Drugs (NTIDs). Should we take the upper limit literally and not based on \(\small{\Delta=10\%}\)? \[\left\{\theta_1,\theta_2\right\}=\left\{100\,(1\Delta),100/(1\Delta)\right\}=90\%,\dot{1}11.11\%\tag{14}\] If yes, we would have to apply double rounding (of the limits and the CI), which should be avoided according to IEEE 7542008.
Health Canada requires for ‘critical dose drugs’ (i.e., NTIDs) that the confidence interval of AUC lies within \(\small{90.0112.0\%}\).^{21} I always thought that \(\small{100/0.900=\dot{1}11.11\%}\).
On the long run products will be approved with \(\small{\sqrt{90.0\times112.0}\approx 100.4\%}\). When asked, the reply was:A similar story, this time dealing with widening the limits for Highly Variable Drugs / Drug Products.^{22} ^{23} ^{24} Based on \(\small{\Delta=25\%}\): \[\left\{\theta_1,\theta_2\right\}=\left\{100\,(1\Delta),100/(1\Delta)\right\}=75\%,1\dot{3}3.33\%\tag{15}\] Shall we use double rounding again?
All jurisdictions accepting Average Bioequivalence with Expanding Limits (ABEL) for HVD(P)s give the regulatory constant \(\small{k}\) in the expansion formula \[\left\{\theta_1,\theta_2\right\}=100\exp(\mp k\,\cdot s_\text{wR})\tag{16}\] with \(\small{0.760}\). Where does it come from?
Based on the switching \(\small{CV_0=30\%}\) we get \[k=\log_{e}1.25 \Big{/}\sqrt{\log_{e}(CV_0^{2}+1)}\approx 0.7601283\ldots\tag{17}\] Not a ‘nice’ number, right? If we would backtransform \(\small{k=0.760}\), the switching \(\small{CV_0}\) would be \(\small{30.00529\%}\)^{25} and not the \(\small{30\%}\) stated in the guideline.^{26}
Health Canada gives the upper cap of scaling in ABEL with \(\small{CV_\text{wR}\;57.4\%}\) but state also that the expansion is limited to \(\small{\left\{\theta_{\text{s}_1},\theta_{\text{s}_2}\right\}=66.7150.0\%}\).^{27} ‘Nice’ numbers as usual but there’s a contradiction.
With \(\small{CV_\text{wR}\;57.4\%}\) we get \[\eqalign{\tag{18}
s_\text{wR}&=\sqrt{\log_{e}(0.574^2+1)}&\approx
0.5336524\ldots\\
\left\{\theta_{\text{s}_1},\theta_{\text{s}_2}\right\}&=100\exp(\mp0.760\cdot
s_\text{wR})&\approx 66.65929\%,150.0166\%
\eqalign}\] In order to obtain \(\small{66.7150.0\%}\), in the power and
sample size functions of PowerTOST
we decided to use a cap
of \(\small{0.57382}\) instead.
< function(x) {
Ottawa < sqrt(log(x^2 + 1))
swR < 100 * exp(0.760 * swR)  150
U
}< uniroot(Ottawa, interval = c(0.3, 1), tol = 1e9)$root
cap < reg_const(regulator = "HC")
hc cat(sprintf("exact cap = %.13f%%", 100 * cap), "(based on U = 150.0%)\n")
print(hc)
setNames(round(100 * scABEL(CV = hc$CVcap, regulator = "HC"), 1),
c("L ", "U "))
# exact cap = 57.3819952841942% (based on U = 150.0%)
# HC regulatory settings
#  CVswitch = 0.3
#  cap on scABEL if CVw(R) > 0.57382
#  regulatory constant = 0.76
#  pe constraint applied
# L U
# 66.7 150.0
It’s guesswork. In coding the package PowerTOST
we
assumed \(\small{66.7150.0\%}\) to be
more ‘important’ than a cap at exactly \(\small{57.4\%}\).
A similar goody in the FDA’s ReferenceScaled Average Bioequivalence (RSABE).^{28} ^{29} Again we have the switching \(\small{CV_0=30\%}\). But then \[s_0=\sqrt{\log_{e}(CV_0^{2}+1)}\approx 0.2935604\ldots\tag{19}\] I thought that there is a consensus about the classification of HVD(P)s, i.e., \(\small{CV_0\;30\%}\). Not for the FDA, giving the regulatory constant with \(\small{0.294}\), which translates to \(\small{30.04689\%}\).
There is a consensus about \(\small{CV_\text{wR}=30\%}\) classifying drugs as highly variable,^{30} although ones with \(\small{CV_\text{wR}=25\%}\) (\(\small{s_\text{wR}=0.2462207}\)) were considered ‘problematic’.^{31} However, the FDA prefers the ‘nicer’ \(\small{\sigma_\text{w0}\;0.25}\), which translates to \(\small{25.39576\%}\).
Instead of ‘nice’ numbers guidelines should state the formulas, which are not more than elementary maths any eight grader could master. It’s not rocket science. We are not retarded. ^{}
All we need are exact numbers of \(\small{\Delta}\) in
ABE and \(\small{CV_0/\sigma_\text{w0}}\) in
SABE.
Everything else should be calculated and assessed with full numerical
precision.
Helmut Schütz 2024
R
and PowerTOST
GPL 3.0,
klippy
MIT,
pandoc
GPL 2.0.
1^{st} version March 30, 2021. Rendered August 4, 2024 12:00
CEST by rmarkdown via pandoc in 0.21 seconds.
Labes D, Schütz H, Lang B. PowerTOST: Power and Sample Size for (Bio)Equivalence Studies. Package version 1.5.6. 20240318. CRAN.↩︎
Fisher RA. Statistical Methods for Research Workers. Chapter III. Distributions. Edinburgh: Oliver & Boyd; 1925.↩︎
HPB, FDA, USP. International Open Conference on Dissolution, Bioavailability, and Bioequivalence. Toronto. June 15–18, 1992.↩︎
Schuirmann DJ. A Comparison of the Two OneSided Tests Procedure and the Power Approach for Assessing the Equivalence of Average Bioavailability. J Pharmacokin Biopharm. 1987; 15(6): 657–80. doi:10.1007/BF01068419.↩︎
Steinijans VW, Hauschke D, Jonkman JHG. Controversies in Bioequivalence Studies. Clin Pharmacokinet. 1992; 22(4): 247–53. doi:10.2165/0000308819922204000001.↩︎
Berger RL, Hsu JC. Bioequivalence Trials, IntersectionUnion Tests and Equivalence Confidence Sets. Stat Sci. 1996; 11(4): 283–302. JSTOR:2246021.↩︎
Where does it come from? Two stories:
Les Benet told
that there was a poll at the FDA and – essentially based on gut
feeling – the 20% saw the light of day.
I’ve heard another one,
which I like more. Wilfred J. Westlake – one of the pioneers of
BE – was a statistician at
SKF. During a
coffee and cig break (most were smoking in the 70s) he asked his fellows
of the clinical pharmacology department »Which difference in blood
concentration do you consider relevant?« Yep, the 20% were born.↩︎
Levy G, Gibaldi M. Bioavailability of Drugs. Circulation. 1974; 49(3): 391–394. doi:10.1161/01.CIR.49.3.391. Open Access.↩︎
Skelly JP. A History of Biopharmaceutics in the Food and Drug Administration 1968–1993. AAPS J. 2010; 12(1): 44–50. doi:10.1208/s1224800991548. Free Full Text.↩︎
FDA, CDER. NDA 204412. Clinical Pharmacology and Biopharmaceutics Review(s). Reference ID: 3244307. Online.↩︎
AAPS, FDA, FIP, HPB, AOAC. Analytical Methods Validation: Bioavailability, Bioequivalence and Pharmacokinetic Studies. Arlington, VA. December 3–5, 1990.↩︎
Think about paracetamol/acetaminophen. With its molecular mass 151.165 g·mol^{–1} and \(\small{f}\) 0.75 after a 500 mg oral dose we start with ~4.94·10^{21} molecules in the circulation. Given its half life of 2½ hours after one week ≈35 molecules still happily float around.↩︎
The declared content \(\small{\neq}\) the measured one!
Furthermore, we can never be sure that the measured content is
the true one. Don’t forget analytical (in)accuracy and
(im)precision. BTW, our method was validated for the test product and
not for the reference. No big deal for simple
IR products. Already more
difficult for MR. Can you hear the
desperate cries of analysts trying to extract the
API from creams
and ointments full of emulsifiers? A dosecorrection is only acceptable
under certain conditions.
The fact that clearances are not identical inflates the
CI, especially for
HVDs (see also this article).↩︎
Possibly geeks like Sheldon Cooper
are fine with \(\small{0.05129329}\).
For most people
\(\small{{\color{Blue}
{95\%}}=100\exp{(0.05129329)}}\) is easier to interpret.↩︎
McGilveray IJ, Dighe SV, French IW, Midha KK (eds). BioInternational ’89. Issues in the Evaluation of Bioavailability Data. Toronto. October 1–4, 1989.↩︎
Keene ON. The log transformation is special. Stat Med. 1995; 14(8): 811–9. doi:10.1002/sim.4780140810. Open Access.↩︎
Mantel N. Do We Want Confidence Intervals Symmetrical About the Null Value? Biometrics. 1977; 33(4): 759–60.↩︎
Kirkwood TBL. Bioequivalence Testing — A Need to Rethink. Biometrics. 1981; 37(3): 589–91. doi:10.2307/2530573.↩︎
I can imagine what people would say:
»Of course, we do. It’s written in the holy books and the
probability of passing increases – that’s great! If regulators don’t
care about an inflated Type I Error, why should we?«↩︎
Goldberg D. What Every Computer Scientist Should Know About FloatingPoint Arithmetic. ACM Computing Surveys. 1991; 23(1): 5–48. doi:10.1145/103162.103163. Open Access.↩︎
Health Canada. Guidance Document. Comparative Bioavailability Standards: Formulations Used for Systemic Effects. Ottawa. 2018/06/08. Online.↩︎
MCC. Registration of Medicines. Biostudies. Pretoria. June 2015. Online.↩︎
GCC. Executive Directorate Of Benefits And Risks Evaluation. The GCC Guidelines for Bioequivalence. Version 2.4. 30/3/2016. Internet Archive.↩︎
League of Arab States, Higher Technical Committee for Arab Pharmaceutical Industry. Harmonised Arab Guideline on Bioequivalence of Generic Pharmaceutical Products. Cairo. March 2014. Online.↩︎
Karalis V, Symillides M, Macheras P. On the levelingoff properties of the new bioequivalence limits for highly variable drugs of the EMA guideline. Eur J Pharm Sci. 2011; 44: 497–505. doi:10.1016/j.ejps.2011.09.008.↩︎
EMA, CHMP. Guideline on the Investigation of Bioequivalence. London. 20 January 2010. Online.↩︎
Health Canada, Therapeutic Products Directorate. Notice: Policy on Bioequivalence Standards for Highly Variable Drug Products. Ottawa. April 18, 2016. Online.↩︎
FDA, OGD. Draft Guidance on Progesterone. Recommended Apr 2010, Revised Feb 2011. Download.↩︎
FDA, CDER. Draft Guidance for Industry. Bioequivalence Studies with Pharmacokinetic Endpoints for Drugs Submitted Under an ANDA. Silver Spring. August 2021. Download.↩︎
McGilveray IJ, Midha KK, Skelly J, Dighe S, Doluisio JT, French IW, Karim A, Burford R. Consensus Report from “Bio International ’89”: Issues in the Evaluation of Bioavailability Data. J Pharm Sci. 1990, 79(10): 945–6. doi:10.1002/jps.2600791022.↩︎
Blume HH, Midha KK. Conference Report. BioInternational 92. Conference on Bioavailability, Bioequivalence and Pharmacokinetic Studies. Bad Homburg, Germany, May 20–22, 1992. In: Midha KK, Blume HH (eds). BioInternational. Bioavailability, Bioequivalence and Pharmacokinetics. Stuttgart: medpharm; 1993.↩︎